diff --git "a/env-llmeval/lib/python3.10/site-packages/scipy/stats/_morestats.py" "b/env-llmeval/lib/python3.10/site-packages/scipy/stats/_morestats.py" new file mode 100644--- /dev/null +++ "b/env-llmeval/lib/python3.10/site-packages/scipy/stats/_morestats.py" @@ -0,0 +1,4933 @@ +from __future__ import annotations +import math +import warnings +from collections import namedtuple + +import numpy as np +from numpy import (isscalar, r_, log, around, unique, asarray, zeros, + arange, sort, amin, amax, sqrt, array, atleast_1d, # noqa: F401 + compress, pi, exp, ravel, count_nonzero, sin, cos, # noqa: F401 + arctan2, hypot) + +from scipy import optimize, special, interpolate, stats +from scipy._lib._bunch import _make_tuple_bunch +from scipy._lib._util import _rename_parameter, _contains_nan, _get_nan + +from ._ansari_swilk_statistics import gscale, swilk +from . import _stats_py, _wilcoxon +from ._fit import FitResult +from ._stats_py import find_repeats, _get_pvalue, SignificanceResult # noqa: F401 +from .contingency import chi2_contingency +from . import distributions +from ._distn_infrastructure import rv_generic +from ._axis_nan_policy import _axis_nan_policy_factory + + +__all__ = ['mvsdist', + 'bayes_mvs', 'kstat', 'kstatvar', 'probplot', 'ppcc_max', 'ppcc_plot', + 'boxcox_llf', 'boxcox', 'boxcox_normmax', 'boxcox_normplot', + 'shapiro', 'anderson', 'ansari', 'bartlett', 'levene', + 'fligner', 'mood', 'wilcoxon', 'median_test', + 'circmean', 'circvar', 'circstd', 'anderson_ksamp', + 'yeojohnson_llf', 'yeojohnson', 'yeojohnson_normmax', + 'yeojohnson_normplot', 'directional_stats', + 'false_discovery_control' + ] + + +Mean = namedtuple('Mean', ('statistic', 'minmax')) +Variance = namedtuple('Variance', ('statistic', 'minmax')) +Std_dev = namedtuple('Std_dev', ('statistic', 'minmax')) + + +def bayes_mvs(data, alpha=0.90): + r""" + Bayesian confidence intervals for the mean, var, and std. + + Parameters + ---------- + data : array_like + Input data, if multi-dimensional it is flattened to 1-D by `bayes_mvs`. + Requires 2 or more data points. + alpha : float, optional + Probability that the returned confidence interval contains + the true parameter. + + Returns + ------- + mean_cntr, var_cntr, std_cntr : tuple + The three results are for the mean, variance and standard deviation, + respectively. Each result is a tuple of the form:: + + (center, (lower, upper)) + + with `center` the mean of the conditional pdf of the value given the + data, and `(lower, upper)` a confidence interval, centered on the + median, containing the estimate to a probability ``alpha``. + + See Also + -------- + mvsdist + + Notes + ----- + Each tuple of mean, variance, and standard deviation estimates represent + the (center, (lower, upper)) with center the mean of the conditional pdf + of the value given the data and (lower, upper) is a confidence interval + centered on the median, containing the estimate to a probability + ``alpha``. + + Converts data to 1-D and assumes all data has the same mean and variance. + Uses Jeffrey's prior for variance and std. + + Equivalent to ``tuple((x.mean(), x.interval(alpha)) for x in mvsdist(dat))`` + + References + ---------- + T.E. Oliphant, "A Bayesian perspective on estimating mean, variance, and + standard-deviation from data", https://scholarsarchive.byu.edu/facpub/278, + 2006. + + Examples + -------- + First a basic example to demonstrate the outputs: + + >>> from scipy import stats + >>> data = [6, 9, 12, 7, 8, 8, 13] + >>> mean, var, std = stats.bayes_mvs(data) + >>> mean + Mean(statistic=9.0, minmax=(7.103650222612533, 10.896349777387467)) + >>> var + Variance(statistic=10.0, minmax=(3.176724206..., 24.45910382...)) + >>> std + Std_dev(statistic=2.9724954732045084, + minmax=(1.7823367265645143, 4.945614605014631)) + + Now we generate some normally distributed random data, and get estimates of + mean and standard deviation with 95% confidence intervals for those + estimates: + + >>> n_samples = 100000 + >>> data = stats.norm.rvs(size=n_samples) + >>> res_mean, res_var, res_std = stats.bayes_mvs(data, alpha=0.95) + + >>> import matplotlib.pyplot as plt + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> ax.hist(data, bins=100, density=True, label='Histogram of data') + >>> ax.vlines(res_mean.statistic, 0, 0.5, colors='r', label='Estimated mean') + >>> ax.axvspan(res_mean.minmax[0],res_mean.minmax[1], facecolor='r', + ... alpha=0.2, label=r'Estimated mean (95% limits)') + >>> ax.vlines(res_std.statistic, 0, 0.5, colors='g', label='Estimated scale') + >>> ax.axvspan(res_std.minmax[0],res_std.minmax[1], facecolor='g', alpha=0.2, + ... label=r'Estimated scale (95% limits)') + + >>> ax.legend(fontsize=10) + >>> ax.set_xlim([-4, 4]) + >>> ax.set_ylim([0, 0.5]) + >>> plt.show() + + """ + m, v, s = mvsdist(data) + if alpha >= 1 or alpha <= 0: + raise ValueError("0 < alpha < 1 is required, but alpha=%s was given." + % alpha) + + m_res = Mean(m.mean(), m.interval(alpha)) + v_res = Variance(v.mean(), v.interval(alpha)) + s_res = Std_dev(s.mean(), s.interval(alpha)) + + return m_res, v_res, s_res + + +def mvsdist(data): + """ + 'Frozen' distributions for mean, variance, and standard deviation of data. + + Parameters + ---------- + data : array_like + Input array. Converted to 1-D using ravel. + Requires 2 or more data-points. + + Returns + ------- + mdist : "frozen" distribution object + Distribution object representing the mean of the data. + vdist : "frozen" distribution object + Distribution object representing the variance of the data. + sdist : "frozen" distribution object + Distribution object representing the standard deviation of the data. + + See Also + -------- + bayes_mvs + + Notes + ----- + The return values from ``bayes_mvs(data)`` is equivalent to + ``tuple((x.mean(), x.interval(0.90)) for x in mvsdist(data))``. + + In other words, calling ``.mean()`` and ``.interval(0.90)`` + on the three distribution objects returned from this function will give + the same results that are returned from `bayes_mvs`. + + References + ---------- + T.E. Oliphant, "A Bayesian perspective on estimating mean, variance, and + standard-deviation from data", https://scholarsarchive.byu.edu/facpub/278, + 2006. + + Examples + -------- + >>> from scipy import stats + >>> data = [6, 9, 12, 7, 8, 8, 13] + >>> mean, var, std = stats.mvsdist(data) + + We now have frozen distribution objects "mean", "var" and "std" that we can + examine: + + >>> mean.mean() + 9.0 + >>> mean.interval(0.95) + (6.6120585482655692, 11.387941451734431) + >>> mean.std() + 1.1952286093343936 + + """ + x = ravel(data) + n = len(x) + if n < 2: + raise ValueError("Need at least 2 data-points.") + xbar = x.mean() + C = x.var() + if n > 1000: # gaussian approximations for large n + mdist = distributions.norm(loc=xbar, scale=math.sqrt(C / n)) + sdist = distributions.norm(loc=math.sqrt(C), scale=math.sqrt(C / (2. * n))) + vdist = distributions.norm(loc=C, scale=math.sqrt(2.0 / n) * C) + else: + nm1 = n - 1 + fac = n * C / 2. + val = nm1 / 2. + mdist = distributions.t(nm1, loc=xbar, scale=math.sqrt(C / nm1)) + sdist = distributions.gengamma(val, -2, scale=math.sqrt(fac)) + vdist = distributions.invgamma(val, scale=fac) + return mdist, vdist, sdist + + +@_axis_nan_policy_factory( + lambda x: x, result_to_tuple=lambda x: (x,), n_outputs=1, default_axis=None +) +def kstat(data, n=2): + r""" + Return the nth k-statistic (1<=n<=4 so far). + + The nth k-statistic k_n is the unique symmetric unbiased estimator of the + nth cumulant kappa_n. + + Parameters + ---------- + data : array_like + Input array. Note that n-D input gets flattened. + n : int, {1, 2, 3, 4}, optional + Default is equal to 2. + + Returns + ------- + kstat : float + The nth k-statistic. + + See Also + -------- + kstatvar : Returns an unbiased estimator of the variance of the k-statistic + moment : Returns the n-th central moment about the mean for a sample. + + Notes + ----- + For a sample size n, the first few k-statistics are given by: + + .. math:: + + k_{1} = \mu + k_{2} = \frac{n}{n-1} m_{2} + k_{3} = \frac{ n^{2} } {(n-1) (n-2)} m_{3} + k_{4} = \frac{ n^{2} [(n + 1)m_{4} - 3(n - 1) m^2_{2}]} {(n-1) (n-2) (n-3)} + + where :math:`\mu` is the sample mean, :math:`m_2` is the sample + variance, and :math:`m_i` is the i-th sample central moment. + + References + ---------- + http://mathworld.wolfram.com/k-Statistic.html + + http://mathworld.wolfram.com/Cumulant.html + + Examples + -------- + >>> from scipy import stats + >>> from numpy.random import default_rng + >>> rng = default_rng() + + As sample size increases, n-th moment and n-th k-statistic converge to the + same number (although they aren't identical). In the case of the normal + distribution, they converge to zero. + + >>> for n in [2, 3, 4, 5, 6, 7]: + ... x = rng.normal(size=10**n) + ... m, k = stats.moment(x, 3), stats.kstat(x, 3) + ... print("%.3g %.3g %.3g" % (m, k, m-k)) + -0.631 -0.651 0.0194 # random + 0.0282 0.0283 -8.49e-05 + -0.0454 -0.0454 1.36e-05 + 7.53e-05 7.53e-05 -2.26e-09 + 0.00166 0.00166 -4.99e-09 + -2.88e-06 -2.88e-06 8.63e-13 + """ + if n > 4 or n < 1: + raise ValueError("k-statistics only supported for 1<=n<=4") + n = int(n) + S = np.zeros(n + 1, np.float64) + data = ravel(data) + N = data.size + + # raise ValueError on empty input + if N == 0: + raise ValueError("Data input must not be empty") + + # on nan input, return nan without warning + if np.isnan(np.sum(data)): + return np.nan + + for k in range(1, n + 1): + S[k] = np.sum(data**k, axis=0) + if n == 1: + return S[1] * 1.0/N + elif n == 2: + return (N*S[2] - S[1]**2.0) / (N*(N - 1.0)) + elif n == 3: + return (2*S[1]**3 - 3*N*S[1]*S[2] + N*N*S[3]) / (N*(N - 1.0)*(N - 2.0)) + elif n == 4: + return ((-6*S[1]**4 + 12*N*S[1]**2 * S[2] - 3*N*(N-1.0)*S[2]**2 - + 4*N*(N+1)*S[1]*S[3] + N*N*(N+1)*S[4]) / + (N*(N-1.0)*(N-2.0)*(N-3.0))) + else: + raise ValueError("Should not be here.") + + +@_axis_nan_policy_factory( + lambda x: x, result_to_tuple=lambda x: (x,), n_outputs=1, default_axis=None +) +def kstatvar(data, n=2): + r"""Return an unbiased estimator of the variance of the k-statistic. + + See `kstat` for more details of the k-statistic. + + Parameters + ---------- + data : array_like + Input array. Note that n-D input gets flattened. + n : int, {1, 2}, optional + Default is equal to 2. + + Returns + ------- + kstatvar : float + The nth k-statistic variance. + + See Also + -------- + kstat : Returns the n-th k-statistic. + moment : Returns the n-th central moment about the mean for a sample. + + Notes + ----- + The variances of the first few k-statistics are given by: + + .. math:: + + var(k_{1}) = \frac{\kappa^2}{n} + var(k_{2}) = \frac{\kappa^4}{n} + \frac{2\kappa^2_{2}}{n - 1} + var(k_{3}) = \frac{\kappa^6}{n} + \frac{9 \kappa_2 \kappa_4}{n - 1} + + \frac{9 \kappa^2_{3}}{n - 1} + + \frac{6 n \kappa^3_{2}}{(n-1) (n-2)} + var(k_{4}) = \frac{\kappa^8}{n} + \frac{16 \kappa_2 \kappa_6}{n - 1} + + \frac{48 \kappa_{3} \kappa_5}{n - 1} + + \frac{34 \kappa^2_{4}}{n-1} + + \frac{72 n \kappa^2_{2} \kappa_4}{(n - 1) (n - 2)} + + \frac{144 n \kappa_{2} \kappa^2_{3}}{(n - 1) (n - 2)} + + \frac{24 (n + 1) n \kappa^4_{2}}{(n - 1) (n - 2) (n - 3)} + """ # noqa: E501 + data = ravel(data) + N = len(data) + if n == 1: + return kstat(data, n=2) * 1.0/N + elif n == 2: + k2 = kstat(data, n=2) + k4 = kstat(data, n=4) + return (2*N*k2**2 + (N-1)*k4) / (N*(N+1)) + else: + raise ValueError("Only n=1 or n=2 supported.") + + +def _calc_uniform_order_statistic_medians(n): + """Approximations of uniform order statistic medians. + + Parameters + ---------- + n : int + Sample size. + + Returns + ------- + v : 1d float array + Approximations of the order statistic medians. + + References + ---------- + .. [1] James J. Filliben, "The Probability Plot Correlation Coefficient + Test for Normality", Technometrics, Vol. 17, pp. 111-117, 1975. + + Examples + -------- + Order statistics of the uniform distribution on the unit interval + are marginally distributed according to beta distributions. + The expectations of these order statistic are evenly spaced across + the interval, but the distributions are skewed in a way that + pushes the medians slightly towards the endpoints of the unit interval: + + >>> import numpy as np + >>> n = 4 + >>> k = np.arange(1, n+1) + >>> from scipy.stats import beta + >>> a = k + >>> b = n-k+1 + >>> beta.mean(a, b) + array([0.2, 0.4, 0.6, 0.8]) + >>> beta.median(a, b) + array([0.15910358, 0.38572757, 0.61427243, 0.84089642]) + + The Filliben approximation uses the exact medians of the smallest + and greatest order statistics, and the remaining medians are approximated + by points spread evenly across a sub-interval of the unit interval: + + >>> from scipy.stats._morestats import _calc_uniform_order_statistic_medians + >>> _calc_uniform_order_statistic_medians(n) + array([0.15910358, 0.38545246, 0.61454754, 0.84089642]) + + This plot shows the skewed distributions of the order statistics + of a sample of size four from a uniform distribution on the unit interval: + + >>> import matplotlib.pyplot as plt + >>> x = np.linspace(0.0, 1.0, num=50, endpoint=True) + >>> pdfs = [beta.pdf(x, a[i], b[i]) for i in range(n)] + >>> plt.figure() + >>> plt.plot(x, pdfs[0], x, pdfs[1], x, pdfs[2], x, pdfs[3]) + + """ + v = np.empty(n, dtype=np.float64) + v[-1] = 0.5**(1.0 / n) + v[0] = 1 - v[-1] + i = np.arange(2, n) + v[1:-1] = (i - 0.3175) / (n + 0.365) + return v + + +def _parse_dist_kw(dist, enforce_subclass=True): + """Parse `dist` keyword. + + Parameters + ---------- + dist : str or stats.distributions instance. + Several functions take `dist` as a keyword, hence this utility + function. + enforce_subclass : bool, optional + If True (default), `dist` needs to be a + `_distn_infrastructure.rv_generic` instance. + It can sometimes be useful to set this keyword to False, if a function + wants to accept objects that just look somewhat like such an instance + (for example, they have a ``ppf`` method). + + """ + if isinstance(dist, rv_generic): + pass + elif isinstance(dist, str): + try: + dist = getattr(distributions, dist) + except AttributeError as e: + raise ValueError("%s is not a valid distribution name" % dist) from e + elif enforce_subclass: + msg = ("`dist` should be a stats.distributions instance or a string " + "with the name of such a distribution.") + raise ValueError(msg) + + return dist + + +def _add_axis_labels_title(plot, xlabel, ylabel, title): + """Helper function to add axes labels and a title to stats plots.""" + try: + if hasattr(plot, 'set_title'): + # Matplotlib Axes instance or something that looks like it + plot.set_title(title) + plot.set_xlabel(xlabel) + plot.set_ylabel(ylabel) + else: + # matplotlib.pyplot module + plot.title(title) + plot.xlabel(xlabel) + plot.ylabel(ylabel) + except Exception: + # Not an MPL object or something that looks (enough) like it. + # Don't crash on adding labels or title + pass + + +def probplot(x, sparams=(), dist='norm', fit=True, plot=None, rvalue=False): + """ + Calculate quantiles for a probability plot, and optionally show the plot. + + Generates a probability plot of sample data against the quantiles of a + specified theoretical distribution (the normal distribution by default). + `probplot` optionally calculates a best-fit line for the data and plots the + results using Matplotlib or a given plot function. + + Parameters + ---------- + x : array_like + Sample/response data from which `probplot` creates the plot. + sparams : tuple, optional + Distribution-specific shape parameters (shape parameters plus location + and scale). + dist : str or stats.distributions instance, optional + Distribution or distribution function name. The default is 'norm' for a + normal probability plot. Objects that look enough like a + stats.distributions instance (i.e. they have a ``ppf`` method) are also + accepted. + fit : bool, optional + Fit a least-squares regression (best-fit) line to the sample data if + True (default). + plot : object, optional + If given, plots the quantiles. + If given and `fit` is True, also plots the least squares fit. + `plot` is an object that has to have methods "plot" and "text". + The `matplotlib.pyplot` module or a Matplotlib Axes object can be used, + or a custom object with the same methods. + Default is None, which means that no plot is created. + rvalue : bool, optional + If `plot` is provided and `fit` is True, setting `rvalue` to True + includes the coefficient of determination on the plot. + Default is False. + + Returns + ------- + (osm, osr) : tuple of ndarrays + Tuple of theoretical quantiles (osm, or order statistic medians) and + ordered responses (osr). `osr` is simply sorted input `x`. + For details on how `osm` is calculated see the Notes section. + (slope, intercept, r) : tuple of floats, optional + Tuple containing the result of the least-squares fit, if that is + performed by `probplot`. `r` is the square root of the coefficient of + determination. If ``fit=False`` and ``plot=None``, this tuple is not + returned. + + Notes + ----- + Even if `plot` is given, the figure is not shown or saved by `probplot`; + ``plt.show()`` or ``plt.savefig('figname.png')`` should be used after + calling `probplot`. + + `probplot` generates a probability plot, which should not be confused with + a Q-Q or a P-P plot. Statsmodels has more extensive functionality of this + type, see ``statsmodels.api.ProbPlot``. + + The formula used for the theoretical quantiles (horizontal axis of the + probability plot) is Filliben's estimate:: + + quantiles = dist.ppf(val), for + + 0.5**(1/n), for i = n + val = (i - 0.3175) / (n + 0.365), for i = 2, ..., n-1 + 1 - 0.5**(1/n), for i = 1 + + where ``i`` indicates the i-th ordered value and ``n`` is the total number + of values. + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + >>> nsample = 100 + >>> rng = np.random.default_rng() + + A t distribution with small degrees of freedom: + + >>> ax1 = plt.subplot(221) + >>> x = stats.t.rvs(3, size=nsample, random_state=rng) + >>> res = stats.probplot(x, plot=plt) + + A t distribution with larger degrees of freedom: + + >>> ax2 = plt.subplot(222) + >>> x = stats.t.rvs(25, size=nsample, random_state=rng) + >>> res = stats.probplot(x, plot=plt) + + A mixture of two normal distributions with broadcasting: + + >>> ax3 = plt.subplot(223) + >>> x = stats.norm.rvs(loc=[0,5], scale=[1,1.5], + ... size=(nsample//2,2), random_state=rng).ravel() + >>> res = stats.probplot(x, plot=plt) + + A standard normal distribution: + + >>> ax4 = plt.subplot(224) + >>> x = stats.norm.rvs(loc=0, scale=1, size=nsample, random_state=rng) + >>> res = stats.probplot(x, plot=plt) + + Produce a new figure with a loggamma distribution, using the ``dist`` and + ``sparams`` keywords: + + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> x = stats.loggamma.rvs(c=2.5, size=500, random_state=rng) + >>> res = stats.probplot(x, dist=stats.loggamma, sparams=(2.5,), plot=ax) + >>> ax.set_title("Probplot for loggamma dist with shape parameter 2.5") + + Show the results with Matplotlib: + + >>> plt.show() + + """ + x = np.asarray(x) + if x.size == 0: + if fit: + return (x, x), (np.nan, np.nan, 0.0) + else: + return x, x + + osm_uniform = _calc_uniform_order_statistic_medians(len(x)) + dist = _parse_dist_kw(dist, enforce_subclass=False) + if sparams is None: + sparams = () + if isscalar(sparams): + sparams = (sparams,) + if not isinstance(sparams, tuple): + sparams = tuple(sparams) + + osm = dist.ppf(osm_uniform, *sparams) + osr = sort(x) + if fit: + # perform a linear least squares fit. + slope, intercept, r, prob, _ = _stats_py.linregress(osm, osr) + + if plot is not None: + plot.plot(osm, osr, 'bo') + if fit: + plot.plot(osm, slope*osm + intercept, 'r-') + _add_axis_labels_title(plot, xlabel='Theoretical quantiles', + ylabel='Ordered Values', + title='Probability Plot') + + # Add R^2 value to the plot as text + if fit and rvalue: + xmin = amin(osm) + xmax = amax(osm) + ymin = amin(x) + ymax = amax(x) + posx = xmin + 0.70 * (xmax - xmin) + posy = ymin + 0.01 * (ymax - ymin) + plot.text(posx, posy, "$R^2=%1.4f$" % r**2) + + if fit: + return (osm, osr), (slope, intercept, r) + else: + return osm, osr + + +def ppcc_max(x, brack=(0.0, 1.0), dist='tukeylambda'): + """Calculate the shape parameter that maximizes the PPCC. + + The probability plot correlation coefficient (PPCC) plot can be used + to determine the optimal shape parameter for a one-parameter family + of distributions. ``ppcc_max`` returns the shape parameter that would + maximize the probability plot correlation coefficient for the given + data to a one-parameter family of distributions. + + Parameters + ---------- + x : array_like + Input array. + brack : tuple, optional + Triple (a,b,c) where (a>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + >>> rng = np.random.default_rng() + >>> c = 2.5 + >>> x = stats.weibull_min.rvs(c, scale=4, size=2000, random_state=rng) + + Generate the PPCC plot for this data with the Weibull distribution. + + >>> fig, ax = plt.subplots(figsize=(8, 6)) + >>> res = stats.ppcc_plot(x, c/2, 2*c, dist='weibull_min', plot=ax) + + We calculate the value where the shape should reach its maximum and a + red line is drawn there. The line should coincide with the highest + point in the PPCC graph. + + >>> cmax = stats.ppcc_max(x, brack=(c/2, 2*c), dist='weibull_min') + >>> ax.axvline(cmax, color='r') + >>> plt.show() + + """ + dist = _parse_dist_kw(dist) + osm_uniform = _calc_uniform_order_statistic_medians(len(x)) + osr = sort(x) + + # this function computes the x-axis values of the probability plot + # and computes a linear regression (including the correlation) + # and returns 1-r so that a minimization function maximizes the + # correlation + def tempfunc(shape, mi, yvals, func): + xvals = func(mi, shape) + r, prob = _stats_py.pearsonr(xvals, yvals) + return 1 - r + + return optimize.brent(tempfunc, brack=brack, + args=(osm_uniform, osr, dist.ppf)) + + +def ppcc_plot(x, a, b, dist='tukeylambda', plot=None, N=80): + """Calculate and optionally plot probability plot correlation coefficient. + + The probability plot correlation coefficient (PPCC) plot can be used to + determine the optimal shape parameter for a one-parameter family of + distributions. It cannot be used for distributions without shape + parameters + (like the normal distribution) or with multiple shape parameters. + + By default a Tukey-Lambda distribution (`stats.tukeylambda`) is used. A + Tukey-Lambda PPCC plot interpolates from long-tailed to short-tailed + distributions via an approximately normal one, and is therefore + particularly useful in practice. + + Parameters + ---------- + x : array_like + Input array. + a, b : scalar + Lower and upper bounds of the shape parameter to use. + dist : str or stats.distributions instance, optional + Distribution or distribution function name. Objects that look enough + like a stats.distributions instance (i.e. they have a ``ppf`` method) + are also accepted. The default is ``'tukeylambda'``. + plot : object, optional + If given, plots PPCC against the shape parameter. + `plot` is an object that has to have methods "plot" and "text". + The `matplotlib.pyplot` module or a Matplotlib Axes object can be used, + or a custom object with the same methods. + Default is None, which means that no plot is created. + N : int, optional + Number of points on the horizontal axis (equally distributed from + `a` to `b`). + + Returns + ------- + svals : ndarray + The shape values for which `ppcc` was calculated. + ppcc : ndarray + The calculated probability plot correlation coefficient values. + + See Also + -------- + ppcc_max, probplot, boxcox_normplot, tukeylambda + + References + ---------- + J.J. Filliben, "The Probability Plot Correlation Coefficient Test for + Normality", Technometrics, Vol. 17, pp. 111-117, 1975. + + Examples + -------- + First we generate some random data from a Weibull distribution + with shape parameter 2.5, and plot the histogram of the data: + + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + >>> rng = np.random.default_rng() + >>> c = 2.5 + >>> x = stats.weibull_min.rvs(c, scale=4, size=2000, random_state=rng) + + Take a look at the histogram of the data. + + >>> fig1, ax = plt.subplots(figsize=(9, 4)) + >>> ax.hist(x, bins=50) + >>> ax.set_title('Histogram of x') + >>> plt.show() + + Now we explore this data with a PPCC plot as well as the related + probability plot and Box-Cox normplot. A red line is drawn where we + expect the PPCC value to be maximal (at the shape parameter ``c`` + used above): + + >>> fig2 = plt.figure(figsize=(12, 4)) + >>> ax1 = fig2.add_subplot(1, 3, 1) + >>> ax2 = fig2.add_subplot(1, 3, 2) + >>> ax3 = fig2.add_subplot(1, 3, 3) + >>> res = stats.probplot(x, plot=ax1) + >>> res = stats.boxcox_normplot(x, -4, 4, plot=ax2) + >>> res = stats.ppcc_plot(x, c/2, 2*c, dist='weibull_min', plot=ax3) + >>> ax3.axvline(c, color='r') + >>> plt.show() + + """ + if b <= a: + raise ValueError("`b` has to be larger than `a`.") + + svals = np.linspace(a, b, num=N) + ppcc = np.empty_like(svals) + for k, sval in enumerate(svals): + _, r2 = probplot(x, sval, dist=dist, fit=True) + ppcc[k] = r2[-1] + + if plot is not None: + plot.plot(svals, ppcc, 'x') + _add_axis_labels_title(plot, xlabel='Shape Values', + ylabel='Prob Plot Corr. Coef.', + title='(%s) PPCC Plot' % dist) + + return svals, ppcc + + +def _log_mean(logx): + # compute log of mean of x from log(x) + return special.logsumexp(logx, axis=0) - np.log(len(logx)) + + +def _log_var(logx): + # compute log of variance of x from log(x) + logmean = _log_mean(logx) + pij = np.full_like(logx, np.pi * 1j, dtype=np.complex128) + logxmu = special.logsumexp([logx, logmean + pij], axis=0) + return np.real(special.logsumexp(2 * logxmu, axis=0)) - np.log(len(logx)) + + +def boxcox_llf(lmb, data): + r"""The boxcox log-likelihood function. + + Parameters + ---------- + lmb : scalar + Parameter for Box-Cox transformation. See `boxcox` for details. + data : array_like + Data to calculate Box-Cox log-likelihood for. If `data` is + multi-dimensional, the log-likelihood is calculated along the first + axis. + + Returns + ------- + llf : float or ndarray + Box-Cox log-likelihood of `data` given `lmb`. A float for 1-D `data`, + an array otherwise. + + See Also + -------- + boxcox, probplot, boxcox_normplot, boxcox_normmax + + Notes + ----- + The Box-Cox log-likelihood function is defined here as + + .. math:: + + llf = (\lambda - 1) \sum_i(\log(x_i)) - + N/2 \log(\sum_i (y_i - \bar{y})^2 / N), + + where ``y`` is the Box-Cox transformed input data ``x``. + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + >>> from mpl_toolkits.axes_grid1.inset_locator import inset_axes + + Generate some random variates and calculate Box-Cox log-likelihood values + for them for a range of ``lmbda`` values: + + >>> rng = np.random.default_rng() + >>> x = stats.loggamma.rvs(5, loc=10, size=1000, random_state=rng) + >>> lmbdas = np.linspace(-2, 10) + >>> llf = np.zeros(lmbdas.shape, dtype=float) + >>> for ii, lmbda in enumerate(lmbdas): + ... llf[ii] = stats.boxcox_llf(lmbda, x) + + Also find the optimal lmbda value with `boxcox`: + + >>> x_most_normal, lmbda_optimal = stats.boxcox(x) + + Plot the log-likelihood as function of lmbda. Add the optimal lmbda as a + horizontal line to check that that's really the optimum: + + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> ax.plot(lmbdas, llf, 'b.-') + >>> ax.axhline(stats.boxcox_llf(lmbda_optimal, x), color='r') + >>> ax.set_xlabel('lmbda parameter') + >>> ax.set_ylabel('Box-Cox log-likelihood') + + Now add some probability plots to show that where the log-likelihood is + maximized the data transformed with `boxcox` looks closest to normal: + + >>> locs = [3, 10, 4] # 'lower left', 'center', 'lower right' + >>> for lmbda, loc in zip([-1, lmbda_optimal, 9], locs): + ... xt = stats.boxcox(x, lmbda=lmbda) + ... (osm, osr), (slope, intercept, r_sq) = stats.probplot(xt) + ... ax_inset = inset_axes(ax, width="20%", height="20%", loc=loc) + ... ax_inset.plot(osm, osr, 'c.', osm, slope*osm + intercept, 'k-') + ... ax_inset.set_xticklabels([]) + ... ax_inset.set_yticklabels([]) + ... ax_inset.set_title(r'$\lambda=%1.2f$' % lmbda) + + >>> plt.show() + + """ + data = np.asarray(data) + N = data.shape[0] + if N == 0: + return np.nan + + logdata = np.log(data) + + # Compute the variance of the transformed data. + if lmb == 0: + logvar = np.log(np.var(logdata, axis=0)) + else: + # Transform without the constant offset 1/lmb. The offset does + # not affect the variance, and the subtraction of the offset can + # lead to loss of precision. + # Division by lmb can be factored out to enhance numerical stability. + logx = lmb * logdata + logvar = _log_var(logx) - 2 * np.log(abs(lmb)) + + return (lmb - 1) * np.sum(logdata, axis=0) - N/2 * logvar + + +def _boxcox_conf_interval(x, lmax, alpha): + # Need to find the lambda for which + # f(x,lmbda) >= f(x,lmax) - 0.5*chi^2_alpha;1 + fac = 0.5 * distributions.chi2.ppf(1 - alpha, 1) + target = boxcox_llf(lmax, x) - fac + + def rootfunc(lmbda, data, target): + return boxcox_llf(lmbda, data) - target + + # Find positive endpoint of interval in which answer is to be found + newlm = lmax + 0.5 + N = 0 + while (rootfunc(newlm, x, target) > 0.0) and (N < 500): + newlm += 0.1 + N += 1 + + if N == 500: + raise RuntimeError("Could not find endpoint.") + + lmplus = optimize.brentq(rootfunc, lmax, newlm, args=(x, target)) + + # Now find negative interval in the same way + newlm = lmax - 0.5 + N = 0 + while (rootfunc(newlm, x, target) > 0.0) and (N < 500): + newlm -= 0.1 + N += 1 + + if N == 500: + raise RuntimeError("Could not find endpoint.") + + lmminus = optimize.brentq(rootfunc, newlm, lmax, args=(x, target)) + return lmminus, lmplus + + +def boxcox(x, lmbda=None, alpha=None, optimizer=None): + r"""Return a dataset transformed by a Box-Cox power transformation. + + Parameters + ---------- + x : ndarray + Input array to be transformed. + + If `lmbda` is not None, this is an alias of + `scipy.special.boxcox`. + Returns nan if ``x < 0``; returns -inf if ``x == 0 and lmbda < 0``. + + If `lmbda` is None, array must be positive, 1-dimensional, and + non-constant. + + lmbda : scalar, optional + If `lmbda` is None (default), find the value of `lmbda` that maximizes + the log-likelihood function and return it as the second output + argument. + + If `lmbda` is not None, do the transformation for that value. + + alpha : float, optional + If `lmbda` is None and `alpha` is not None (default), return the + ``100 * (1-alpha)%`` confidence interval for `lmbda` as the third + output argument. Must be between 0.0 and 1.0. + + If `lmbda` is not None, `alpha` is ignored. + optimizer : callable, optional + If `lmbda` is None, `optimizer` is the scalar optimizer used to find + the value of `lmbda` that minimizes the negative log-likelihood + function. `optimizer` is a callable that accepts one argument: + + fun : callable + The objective function, which evaluates the negative + log-likelihood function at a provided value of `lmbda` + + and returns an object, such as an instance of + `scipy.optimize.OptimizeResult`, which holds the optimal value of + `lmbda` in an attribute `x`. + + See the example in `boxcox_normmax` or the documentation of + `scipy.optimize.minimize_scalar` for more information. + + If `lmbda` is not None, `optimizer` is ignored. + + Returns + ------- + boxcox : ndarray + Box-Cox power transformed array. + maxlog : float, optional + If the `lmbda` parameter is None, the second returned argument is + the `lmbda` that maximizes the log-likelihood function. + (min_ci, max_ci) : tuple of float, optional + If `lmbda` parameter is None and `alpha` is not None, this returned + tuple of floats represents the minimum and maximum confidence limits + given `alpha`. + + See Also + -------- + probplot, boxcox_normplot, boxcox_normmax, boxcox_llf + + Notes + ----- + The Box-Cox transform is given by:: + + y = (x**lmbda - 1) / lmbda, for lmbda != 0 + log(x), for lmbda = 0 + + `boxcox` requires the input data to be positive. Sometimes a Box-Cox + transformation provides a shift parameter to achieve this; `boxcox` does + not. Such a shift parameter is equivalent to adding a positive constant to + `x` before calling `boxcox`. + + The confidence limits returned when `alpha` is provided give the interval + where: + + .. math:: + + llf(\hat{\lambda}) - llf(\lambda) < \frac{1}{2}\chi^2(1 - \alpha, 1), + + with ``llf`` the log-likelihood function and :math:`\chi^2` the chi-squared + function. + + References + ---------- + G.E.P. Box and D.R. Cox, "An Analysis of Transformations", Journal of the + Royal Statistical Society B, 26, 211-252 (1964). + + Examples + -------- + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + We generate some random variates from a non-normal distribution and make a + probability plot for it, to show it is non-normal in the tails: + + >>> fig = plt.figure() + >>> ax1 = fig.add_subplot(211) + >>> x = stats.loggamma.rvs(5, size=500) + 5 + >>> prob = stats.probplot(x, dist=stats.norm, plot=ax1) + >>> ax1.set_xlabel('') + >>> ax1.set_title('Probplot against normal distribution') + + We now use `boxcox` to transform the data so it's closest to normal: + + >>> ax2 = fig.add_subplot(212) + >>> xt, _ = stats.boxcox(x) + >>> prob = stats.probplot(xt, dist=stats.norm, plot=ax2) + >>> ax2.set_title('Probplot after Box-Cox transformation') + + >>> plt.show() + + """ + x = np.asarray(x) + + if lmbda is not None: # single transformation + return special.boxcox(x, lmbda) + + if x.ndim != 1: + raise ValueError("Data must be 1-dimensional.") + + if x.size == 0: + return x + + if np.all(x == x[0]): + raise ValueError("Data must not be constant.") + + if np.any(x <= 0): + raise ValueError("Data must be positive.") + + # If lmbda=None, find the lmbda that maximizes the log-likelihood function. + lmax = boxcox_normmax(x, method='mle', optimizer=optimizer) + y = boxcox(x, lmax) + + if alpha is None: + return y, lmax + else: + # Find confidence interval + interval = _boxcox_conf_interval(x, lmax, alpha) + return y, lmax, interval + + +def _boxcox_inv_lmbda(x, y): + # compute lmbda given x and y for Box-Cox transformation + num = special.lambertw(-(x ** (-1 / y)) * np.log(x) / y, k=-1) + return np.real(-num / np.log(x) - 1 / y) + + +class _BigFloat: + def __repr__(self): + return "BIG_FLOAT" + + +def boxcox_normmax( + x, brack=None, method='pearsonr', optimizer=None, *, ymax=_BigFloat() +): + """Compute optimal Box-Cox transform parameter for input data. + + Parameters + ---------- + x : array_like + Input array. All entries must be positive, finite, real numbers. + brack : 2-tuple, optional, default (-2.0, 2.0) + The starting interval for a downhill bracket search for the default + `optimize.brent` solver. Note that this is in most cases not + critical; the final result is allowed to be outside this bracket. + If `optimizer` is passed, `brack` must be None. + method : str, optional + The method to determine the optimal transform parameter (`boxcox` + ``lmbda`` parameter). Options are: + + 'pearsonr' (default) + Maximizes the Pearson correlation coefficient between + ``y = boxcox(x)`` and the expected values for ``y`` if `x` would be + normally-distributed. + + 'mle' + Maximizes the log-likelihood `boxcox_llf`. This is the method used + in `boxcox`. + + 'all' + Use all optimization methods available, and return all results. + Useful to compare different methods. + optimizer : callable, optional + `optimizer` is a callable that accepts one argument: + + fun : callable + The objective function to be minimized. `fun` accepts one argument, + the Box-Cox transform parameter `lmbda`, and returns the value of + the function (e.g., the negative log-likelihood) at the provided + argument. The job of `optimizer` is to find the value of `lmbda` + that *minimizes* `fun`. + + and returns an object, such as an instance of + `scipy.optimize.OptimizeResult`, which holds the optimal value of + `lmbda` in an attribute `x`. + + See the example below or the documentation of + `scipy.optimize.minimize_scalar` for more information. + ymax : float, optional + The unconstrained optimal transform parameter may cause Box-Cox + transformed data to have extreme magnitude or even overflow. + This parameter constrains MLE optimization such that the magnitude + of the transformed `x` does not exceed `ymax`. The default is + the maximum value of the input dtype. If set to infinity, + `boxcox_normmax` returns the unconstrained optimal lambda. + Ignored when ``method='pearsonr'``. + + Returns + ------- + maxlog : float or ndarray + The optimal transform parameter found. An array instead of a scalar + for ``method='all'``. + + See Also + -------- + boxcox, boxcox_llf, boxcox_normplot, scipy.optimize.minimize_scalar + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + We can generate some data and determine the optimal ``lmbda`` in various + ways: + + >>> rng = np.random.default_rng() + >>> x = stats.loggamma.rvs(5, size=30, random_state=rng) + 5 + >>> y, lmax_mle = stats.boxcox(x) + >>> lmax_pearsonr = stats.boxcox_normmax(x) + + >>> lmax_mle + 2.217563431465757 + >>> lmax_pearsonr + 2.238318660200961 + >>> stats.boxcox_normmax(x, method='all') + array([2.23831866, 2.21756343]) + + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> prob = stats.boxcox_normplot(x, -10, 10, plot=ax) + >>> ax.axvline(lmax_mle, color='r') + >>> ax.axvline(lmax_pearsonr, color='g', ls='--') + + >>> plt.show() + + Alternatively, we can define our own `optimizer` function. Suppose we + are only interested in values of `lmbda` on the interval [6, 7], we + want to use `scipy.optimize.minimize_scalar` with ``method='bounded'``, + and we want to use tighter tolerances when optimizing the log-likelihood + function. To do this, we define a function that accepts positional argument + `fun` and uses `scipy.optimize.minimize_scalar` to minimize `fun` subject + to the provided bounds and tolerances: + + >>> from scipy import optimize + >>> options = {'xatol': 1e-12} # absolute tolerance on `x` + >>> def optimizer(fun): + ... return optimize.minimize_scalar(fun, bounds=(6, 7), + ... method="bounded", options=options) + >>> stats.boxcox_normmax(x, optimizer=optimizer) + 6.000... + """ + x = np.asarray(x) + end_msg = "exceed specified `ymax`." + if isinstance(ymax, _BigFloat): + dtype = x.dtype if np.issubdtype(x.dtype, np.floating) else np.float64 + # 10000 is a safety factor because `special.boxcox` overflows prematurely. + ymax = np.finfo(dtype).max / 10000 + end_msg = f"overflow in {dtype}." + elif ymax <= 0: + raise ValueError("`ymax` must be strictly positive") + + # If optimizer is not given, define default 'brent' optimizer. + if optimizer is None: + + # Set default value for `brack`. + if brack is None: + brack = (-2.0, 2.0) + + def _optimizer(func, args): + return optimize.brent(func, args=args, brack=brack) + + # Otherwise check optimizer. + else: + if not callable(optimizer): + raise ValueError("`optimizer` must be a callable") + + if brack is not None: + raise ValueError("`brack` must be None if `optimizer` is given") + + # `optimizer` is expected to return a `OptimizeResult` object, we here + # get the solution to the optimization problem. + def _optimizer(func, args): + def func_wrapped(x): + return func(x, *args) + return getattr(optimizer(func_wrapped), 'x', None) + + def _pearsonr(x): + osm_uniform = _calc_uniform_order_statistic_medians(len(x)) + xvals = distributions.norm.ppf(osm_uniform) + + def _eval_pearsonr(lmbda, xvals, samps): + # This function computes the x-axis values of the probability plot + # and computes a linear regression (including the correlation) and + # returns ``1 - r`` so that a minimization function maximizes the + # correlation. + y = boxcox(samps, lmbda) + yvals = np.sort(y) + r, prob = _stats_py.pearsonr(xvals, yvals) + return 1 - r + + return _optimizer(_eval_pearsonr, args=(xvals, x)) + + def _mle(x): + def _eval_mle(lmb, data): + # function to minimize + return -boxcox_llf(lmb, data) + + return _optimizer(_eval_mle, args=(x,)) + + def _all(x): + maxlog = np.empty(2, dtype=float) + maxlog[0] = _pearsonr(x) + maxlog[1] = _mle(x) + return maxlog + + methods = {'pearsonr': _pearsonr, + 'mle': _mle, + 'all': _all} + if method not in methods.keys(): + raise ValueError("Method %s not recognized." % method) + + optimfunc = methods[method] + + try: + res = optimfunc(x) + except ValueError as e: + if "infs or NaNs" in str(e): + message = ("The `x` argument of `boxcox_normmax` must contain " + "only positive, finite, real numbers.") + raise ValueError(message) from e + else: + raise e + + if res is None: + message = ("The `optimizer` argument of `boxcox_normmax` must return " + "an object containing the optimal `lmbda` in attribute `x`.") + raise ValueError(message) + elif not np.isinf(ymax): # adjust the final lambda + # x > 1, boxcox(x) > 0; x < 1, boxcox(x) < 0 + xmax, xmin = np.max(x), np.min(x) + if xmin >= 1: + x_treme = xmax + elif xmax <= 1: + x_treme = xmin + else: # xmin < 1 < xmax + indicator = special.boxcox(xmax, res) > abs(special.boxcox(xmin, res)) + if isinstance(res, np.ndarray): + indicator = indicator[1] # select corresponds with 'mle' + x_treme = xmax if indicator else xmin + + mask = abs(special.boxcox(x_treme, res)) > ymax + if np.any(mask): + message = ( + f"The optimal lambda is {res}, but the returned lambda is the " + f"constrained optimum to ensure that the maximum or the minimum " + f"of the transformed data does not " + end_msg + ) + warnings.warn(message, stacklevel=2) + + # Return the constrained lambda to ensure the transformation + # does not cause overflow or exceed specified `ymax` + constrained_res = _boxcox_inv_lmbda(x_treme, ymax * np.sign(x_treme - 1)) + + if isinstance(res, np.ndarray): + res[mask] = constrained_res + else: + res = constrained_res + return res + + +def _normplot(method, x, la, lb, plot=None, N=80): + """Compute parameters for a Box-Cox or Yeo-Johnson normality plot, + optionally show it. + + See `boxcox_normplot` or `yeojohnson_normplot` for details. + """ + + if method == 'boxcox': + title = 'Box-Cox Normality Plot' + transform_func = boxcox + else: + title = 'Yeo-Johnson Normality Plot' + transform_func = yeojohnson + + x = np.asarray(x) + if x.size == 0: + return x + + if lb <= la: + raise ValueError("`lb` has to be larger than `la`.") + + if method == 'boxcox' and np.any(x <= 0): + raise ValueError("Data must be positive.") + + lmbdas = np.linspace(la, lb, num=N) + ppcc = lmbdas * 0.0 + for i, val in enumerate(lmbdas): + # Determine for each lmbda the square root of correlation coefficient + # of transformed x + z = transform_func(x, lmbda=val) + _, (_, _, r) = probplot(z, dist='norm', fit=True) + ppcc[i] = r + + if plot is not None: + plot.plot(lmbdas, ppcc, 'x') + _add_axis_labels_title(plot, xlabel='$\\lambda$', + ylabel='Prob Plot Corr. Coef.', + title=title) + + return lmbdas, ppcc + + +def boxcox_normplot(x, la, lb, plot=None, N=80): + """Compute parameters for a Box-Cox normality plot, optionally show it. + + A Box-Cox normality plot shows graphically what the best transformation + parameter is to use in `boxcox` to obtain a distribution that is close + to normal. + + Parameters + ---------- + x : array_like + Input array. + la, lb : scalar + The lower and upper bounds for the ``lmbda`` values to pass to `boxcox` + for Box-Cox transformations. These are also the limits of the + horizontal axis of the plot if that is generated. + plot : object, optional + If given, plots the quantiles and least squares fit. + `plot` is an object that has to have methods "plot" and "text". + The `matplotlib.pyplot` module or a Matplotlib Axes object can be used, + or a custom object with the same methods. + Default is None, which means that no plot is created. + N : int, optional + Number of points on the horizontal axis (equally distributed from + `la` to `lb`). + + Returns + ------- + lmbdas : ndarray + The ``lmbda`` values for which a Box-Cox transform was done. + ppcc : ndarray + Probability Plot Correlelation Coefficient, as obtained from `probplot` + when fitting the Box-Cox transformed input `x` against a normal + distribution. + + See Also + -------- + probplot, boxcox, boxcox_normmax, boxcox_llf, ppcc_max + + Notes + ----- + Even if `plot` is given, the figure is not shown or saved by + `boxcox_normplot`; ``plt.show()`` or ``plt.savefig('figname.png')`` + should be used after calling `probplot`. + + Examples + -------- + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + Generate some non-normally distributed data, and create a Box-Cox plot: + + >>> x = stats.loggamma.rvs(5, size=500) + 5 + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> prob = stats.boxcox_normplot(x, -20, 20, plot=ax) + + Determine and plot the optimal ``lmbda`` to transform ``x`` and plot it in + the same plot: + + >>> _, maxlog = stats.boxcox(x) + >>> ax.axvline(maxlog, color='r') + + >>> plt.show() + + """ + return _normplot('boxcox', x, la, lb, plot, N) + + +def yeojohnson(x, lmbda=None): + r"""Return a dataset transformed by a Yeo-Johnson power transformation. + + Parameters + ---------- + x : ndarray + Input array. Should be 1-dimensional. + lmbda : float, optional + If ``lmbda`` is ``None``, find the lambda that maximizes the + log-likelihood function and return it as the second output argument. + Otherwise the transformation is done for the given value. + + Returns + ------- + yeojohnson: ndarray + Yeo-Johnson power transformed array. + maxlog : float, optional + If the `lmbda` parameter is None, the second returned argument is + the lambda that maximizes the log-likelihood function. + + See Also + -------- + probplot, yeojohnson_normplot, yeojohnson_normmax, yeojohnson_llf, boxcox + + Notes + ----- + The Yeo-Johnson transform is given by:: + + y = ((x + 1)**lmbda - 1) / lmbda, for x >= 0, lmbda != 0 + log(x + 1), for x >= 0, lmbda = 0 + -((-x + 1)**(2 - lmbda) - 1) / (2 - lmbda), for x < 0, lmbda != 2 + -log(-x + 1), for x < 0, lmbda = 2 + + Unlike `boxcox`, `yeojohnson` does not require the input data to be + positive. + + .. versionadded:: 1.2.0 + + + References + ---------- + I. Yeo and R.A. Johnson, "A New Family of Power Transformations to + Improve Normality or Symmetry", Biometrika 87.4 (2000): + + + Examples + -------- + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + We generate some random variates from a non-normal distribution and make a + probability plot for it, to show it is non-normal in the tails: + + >>> fig = plt.figure() + >>> ax1 = fig.add_subplot(211) + >>> x = stats.loggamma.rvs(5, size=500) + 5 + >>> prob = stats.probplot(x, dist=stats.norm, plot=ax1) + >>> ax1.set_xlabel('') + >>> ax1.set_title('Probplot against normal distribution') + + We now use `yeojohnson` to transform the data so it's closest to normal: + + >>> ax2 = fig.add_subplot(212) + >>> xt, lmbda = stats.yeojohnson(x) + >>> prob = stats.probplot(xt, dist=stats.norm, plot=ax2) + >>> ax2.set_title('Probplot after Yeo-Johnson transformation') + + >>> plt.show() + + """ + x = np.asarray(x) + if x.size == 0: + return x + + if np.issubdtype(x.dtype, np.complexfloating): + raise ValueError('Yeo-Johnson transformation is not defined for ' + 'complex numbers.') + + if np.issubdtype(x.dtype, np.integer): + x = x.astype(np.float64, copy=False) + + if lmbda is not None: + return _yeojohnson_transform(x, lmbda) + + # if lmbda=None, find the lmbda that maximizes the log-likelihood function. + lmax = yeojohnson_normmax(x) + y = _yeojohnson_transform(x, lmax) + + return y, lmax + + +def _yeojohnson_transform(x, lmbda): + """Returns `x` transformed by the Yeo-Johnson power transform with given + parameter `lmbda`. + """ + dtype = x.dtype if np.issubdtype(x.dtype, np.floating) else np.float64 + out = np.zeros_like(x, dtype=dtype) + pos = x >= 0 # binary mask + + # when x >= 0 + if abs(lmbda) < np.spacing(1.): + out[pos] = np.log1p(x[pos]) + else: # lmbda != 0 + # more stable version of: ((x + 1) ** lmbda - 1) / lmbda + out[pos] = np.expm1(lmbda * np.log1p(x[pos])) / lmbda + + # when x < 0 + if abs(lmbda - 2) > np.spacing(1.): + out[~pos] = -np.expm1((2 - lmbda) * np.log1p(-x[~pos])) / (2 - lmbda) + else: # lmbda == 2 + out[~pos] = -np.log1p(-x[~pos]) + + return out + + +def yeojohnson_llf(lmb, data): + r"""The yeojohnson log-likelihood function. + + Parameters + ---------- + lmb : scalar + Parameter for Yeo-Johnson transformation. See `yeojohnson` for + details. + data : array_like + Data to calculate Yeo-Johnson log-likelihood for. If `data` is + multi-dimensional, the log-likelihood is calculated along the first + axis. + + Returns + ------- + llf : float + Yeo-Johnson log-likelihood of `data` given `lmb`. + + See Also + -------- + yeojohnson, probplot, yeojohnson_normplot, yeojohnson_normmax + + Notes + ----- + The Yeo-Johnson log-likelihood function is defined here as + + .. math:: + + llf = -N/2 \log(\hat{\sigma}^2) + (\lambda - 1) + \sum_i \text{ sign }(x_i)\log(|x_i| + 1) + + where :math:`\hat{\sigma}^2` is estimated variance of the Yeo-Johnson + transformed input data ``x``. + + .. versionadded:: 1.2.0 + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + >>> from mpl_toolkits.axes_grid1.inset_locator import inset_axes + + Generate some random variates and calculate Yeo-Johnson log-likelihood + values for them for a range of ``lmbda`` values: + + >>> x = stats.loggamma.rvs(5, loc=10, size=1000) + >>> lmbdas = np.linspace(-2, 10) + >>> llf = np.zeros(lmbdas.shape, dtype=float) + >>> for ii, lmbda in enumerate(lmbdas): + ... llf[ii] = stats.yeojohnson_llf(lmbda, x) + + Also find the optimal lmbda value with `yeojohnson`: + + >>> x_most_normal, lmbda_optimal = stats.yeojohnson(x) + + Plot the log-likelihood as function of lmbda. Add the optimal lmbda as a + horizontal line to check that that's really the optimum: + + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> ax.plot(lmbdas, llf, 'b.-') + >>> ax.axhline(stats.yeojohnson_llf(lmbda_optimal, x), color='r') + >>> ax.set_xlabel('lmbda parameter') + >>> ax.set_ylabel('Yeo-Johnson log-likelihood') + + Now add some probability plots to show that where the log-likelihood is + maximized the data transformed with `yeojohnson` looks closest to normal: + + >>> locs = [3, 10, 4] # 'lower left', 'center', 'lower right' + >>> for lmbda, loc in zip([-1, lmbda_optimal, 9], locs): + ... xt = stats.yeojohnson(x, lmbda=lmbda) + ... (osm, osr), (slope, intercept, r_sq) = stats.probplot(xt) + ... ax_inset = inset_axes(ax, width="20%", height="20%", loc=loc) + ... ax_inset.plot(osm, osr, 'c.', osm, slope*osm + intercept, 'k-') + ... ax_inset.set_xticklabels([]) + ... ax_inset.set_yticklabels([]) + ... ax_inset.set_title(r'$\lambda=%1.2f$' % lmbda) + + >>> plt.show() + + """ + data = np.asarray(data) + n_samples = data.shape[0] + + if n_samples == 0: + return np.nan + + trans = _yeojohnson_transform(data, lmb) + trans_var = trans.var(axis=0) + loglike = np.empty_like(trans_var) + + # Avoid RuntimeWarning raised by np.log when the variance is too low + tiny_variance = trans_var < np.finfo(trans_var.dtype).tiny + loglike[tiny_variance] = np.inf + + loglike[~tiny_variance] = ( + -n_samples / 2 * np.log(trans_var[~tiny_variance])) + loglike[~tiny_variance] += ( + (lmb - 1) * (np.sign(data) * np.log1p(np.abs(data))).sum(axis=0)) + return loglike + + +def yeojohnson_normmax(x, brack=None): + """Compute optimal Yeo-Johnson transform parameter. + + Compute optimal Yeo-Johnson transform parameter for input data, using + maximum likelihood estimation. + + Parameters + ---------- + x : array_like + Input array. + brack : 2-tuple, optional + The starting interval for a downhill bracket search with + `optimize.brent`. Note that this is in most cases not critical; the + final result is allowed to be outside this bracket. If None, + `optimize.fminbound` is used with bounds that avoid overflow. + + Returns + ------- + maxlog : float + The optimal transform parameter found. + + See Also + -------- + yeojohnson, yeojohnson_llf, yeojohnson_normplot + + Notes + ----- + .. versionadded:: 1.2.0 + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + Generate some data and determine optimal ``lmbda`` + + >>> rng = np.random.default_rng() + >>> x = stats.loggamma.rvs(5, size=30, random_state=rng) + 5 + >>> lmax = stats.yeojohnson_normmax(x) + + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> prob = stats.yeojohnson_normplot(x, -10, 10, plot=ax) + >>> ax.axvline(lmax, color='r') + + >>> plt.show() + + """ + def _neg_llf(lmbda, data): + llf = yeojohnson_llf(lmbda, data) + # reject likelihoods that are inf which are likely due to small + # variance in the transformed space + llf[np.isinf(llf)] = -np.inf + return -llf + + with np.errstate(invalid='ignore'): + if not np.all(np.isfinite(x)): + raise ValueError('Yeo-Johnson input must be finite.') + if np.all(x == 0): + return 1.0 + if brack is not None: + return optimize.brent(_neg_llf, brack=brack, args=(x,)) + x = np.asarray(x) + dtype = x.dtype if np.issubdtype(x.dtype, np.floating) else np.float64 + # Allow values up to 20 times the maximum observed value to be safely + # transformed without over- or underflow. + log1p_max_x = np.log1p(20 * np.max(np.abs(x))) + # Use half of floating point's exponent range to allow safe computation + # of the variance of the transformed data. + log_eps = np.log(np.finfo(dtype).eps) + log_tiny_float = (np.log(np.finfo(dtype).tiny) - log_eps) / 2 + log_max_float = (np.log(np.finfo(dtype).max) + log_eps) / 2 + # Compute the bounds by approximating the inverse of the Yeo-Johnson + # transform on the smallest and largest floating point exponents, given + # the largest data we expect to observe. See [1] for further details. + # [1] https://github.com/scipy/scipy/pull/18852#issuecomment-1630286174 + lb = log_tiny_float / log1p_max_x + ub = log_max_float / log1p_max_x + # Convert the bounds if all or some of the data is negative. + if np.all(x < 0): + lb, ub = 2 - ub, 2 - lb + elif np.any(x < 0): + lb, ub = max(2 - ub, lb), min(2 - lb, ub) + # Match `optimize.brent`'s tolerance. + tol_brent = 1.48e-08 + return optimize.fminbound(_neg_llf, lb, ub, args=(x,), xtol=tol_brent) + + +def yeojohnson_normplot(x, la, lb, plot=None, N=80): + """Compute parameters for a Yeo-Johnson normality plot, optionally show it. + + A Yeo-Johnson normality plot shows graphically what the best + transformation parameter is to use in `yeojohnson` to obtain a + distribution that is close to normal. + + Parameters + ---------- + x : array_like + Input array. + la, lb : scalar + The lower and upper bounds for the ``lmbda`` values to pass to + `yeojohnson` for Yeo-Johnson transformations. These are also the + limits of the horizontal axis of the plot if that is generated. + plot : object, optional + If given, plots the quantiles and least squares fit. + `plot` is an object that has to have methods "plot" and "text". + The `matplotlib.pyplot` module or a Matplotlib Axes object can be used, + or a custom object with the same methods. + Default is None, which means that no plot is created. + N : int, optional + Number of points on the horizontal axis (equally distributed from + `la` to `lb`). + + Returns + ------- + lmbdas : ndarray + The ``lmbda`` values for which a Yeo-Johnson transform was done. + ppcc : ndarray + Probability Plot Correlelation Coefficient, as obtained from `probplot` + when fitting the Box-Cox transformed input `x` against a normal + distribution. + + See Also + -------- + probplot, yeojohnson, yeojohnson_normmax, yeojohnson_llf, ppcc_max + + Notes + ----- + Even if `plot` is given, the figure is not shown or saved by + `boxcox_normplot`; ``plt.show()`` or ``plt.savefig('figname.png')`` + should be used after calling `probplot`. + + .. versionadded:: 1.2.0 + + Examples + -------- + >>> from scipy import stats + >>> import matplotlib.pyplot as plt + + Generate some non-normally distributed data, and create a Yeo-Johnson plot: + + >>> x = stats.loggamma.rvs(5, size=500) + 5 + >>> fig = plt.figure() + >>> ax = fig.add_subplot(111) + >>> prob = stats.yeojohnson_normplot(x, -20, 20, plot=ax) + + Determine and plot the optimal ``lmbda`` to transform ``x`` and plot it in + the same plot: + + >>> _, maxlog = stats.yeojohnson(x) + >>> ax.axvline(maxlog, color='r') + + >>> plt.show() + + """ + return _normplot('yeojohnson', x, la, lb, plot, N) + + +ShapiroResult = namedtuple('ShapiroResult', ('statistic', 'pvalue')) + + +@_axis_nan_policy_factory(ShapiroResult, n_samples=1, too_small=2, default_axis=None) +def shapiro(x): + r"""Perform the Shapiro-Wilk test for normality. + + The Shapiro-Wilk test tests the null hypothesis that the + data was drawn from a normal distribution. + + Parameters + ---------- + x : array_like + Array of sample data. + + Returns + ------- + statistic : float + The test statistic. + p-value : float + The p-value for the hypothesis test. + + See Also + -------- + anderson : The Anderson-Darling test for normality + kstest : The Kolmogorov-Smirnov test for goodness of fit. + + Notes + ----- + The algorithm used is described in [4]_ but censoring parameters as + described are not implemented. For N > 5000 the W test statistic is + accurate, but the p-value may not be. + + References + ---------- + .. [1] https://www.itl.nist.gov/div898/handbook/prc/section2/prc213.htm + :doi:`10.18434/M32189` + .. [2] Shapiro, S. S. & Wilk, M.B, "An analysis of variance test for + normality (complete samples)", Biometrika, 1965, Vol. 52, + pp. 591-611, :doi:`10.2307/2333709` + .. [3] Razali, N. M. & Wah, Y. B., "Power comparisons of Shapiro-Wilk, + Kolmogorov-Smirnov, Lilliefors and Anderson-Darling tests", Journal + of Statistical Modeling and Analytics, 2011, Vol. 2, pp. 21-33. + .. [4] Royston P., "Remark AS R94: A Remark on Algorithm AS 181: The + W-test for Normality", 1995, Applied Statistics, Vol. 44, + :doi:`10.2307/2986146` + .. [5] Phipson B., and Smyth, G. K., "Permutation P-values Should Never Be + Zero: Calculating Exact P-values When Permutations Are Randomly + Drawn", Statistical Applications in Genetics and Molecular Biology, + 2010, Vol.9, :doi:`10.2202/1544-6115.1585` + .. [6] Panagiotakos, D. B., "The value of p-value in biomedical + research", The Open Cardiovascular Medicine Journal, 2008, Vol.2, + pp. 97-99, :doi:`10.2174/1874192400802010097` + + Examples + -------- + Suppose we wish to infer from measurements whether the weights of adult + human males in a medical study are not normally distributed [2]_. + The weights (lbs) are recorded in the array ``x`` below. + + >>> import numpy as np + >>> x = np.array([148, 154, 158, 160, 161, 162, 166, 170, 182, 195, 236]) + + The normality test of [1]_ and [2]_ begins by computing a statistic based + on the relationship between the observations and the expected order + statistics of a normal distribution. + + >>> from scipy import stats + >>> res = stats.shapiro(x) + >>> res.statistic + 0.7888147830963135 + + The value of this statistic tends to be high (close to 1) for samples drawn + from a normal distribution. + + The test is performed by comparing the observed value of the statistic + against the null distribution: the distribution of statistic values formed + under the null hypothesis that the weights were drawn from a normal + distribution. For this normality test, the null distribution is not easy to + calculate exactly, so it is usually approximated by Monte Carlo methods, + that is, drawing many samples of the same size as ``x`` from a normal + distribution and computing the values of the statistic for each. + + >>> def statistic(x): + ... # Get only the `shapiro` statistic; ignore its p-value + ... return stats.shapiro(x).statistic + >>> ref = stats.monte_carlo_test(x, stats.norm.rvs, statistic, + ... alternative='less') + >>> import matplotlib.pyplot as plt + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> bins = np.linspace(0.65, 1, 50) + >>> def plot(ax): # we'll reuse this + ... ax.hist(ref.null_distribution, density=True, bins=bins) + ... ax.set_title("Shapiro-Wilk Test Null Distribution \n" + ... "(Monte Carlo Approximation, 11 Observations)") + ... ax.set_xlabel("statistic") + ... ax.set_ylabel("probability density") + >>> plot(ax) + >>> plt.show() + + The comparison is quantified by the p-value: the proportion of values in + the null distribution less than or equal to the observed value of the + statistic. + + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> annotation = (f'p-value={res.pvalue:.6f}\n(highlighted area)') + >>> props = dict(facecolor='black', width=1, headwidth=5, headlength=8) + >>> _ = ax.annotate(annotation, (0.75, 0.1), (0.68, 0.7), arrowprops=props) + >>> i_extreme = np.where(bins <= res.statistic)[0] + >>> for i in i_extreme: + ... ax.patches[i].set_color('C1') + >>> plt.xlim(0.65, 0.9) + >>> plt.ylim(0, 4) + >>> plt.show + >>> res.pvalue + 0.006703833118081093 + + If the p-value is "small" - that is, if there is a low probability of + sampling data from a normally distributed population that produces such an + extreme value of the statistic - this may be taken as evidence against + the null hypothesis in favor of the alternative: the weights were not + drawn from a normal distribution. Note that: + + - The inverse is not true; that is, the test is not used to provide + evidence *for* the null hypothesis. + - The threshold for values that will be considered "small" is a choice that + should be made before the data is analyzed [5]_ with consideration of the + risks of both false positives (incorrectly rejecting the null hypothesis) + and false negatives (failure to reject a false null hypothesis). + + """ + x = np.ravel(x).astype(np.float64) + + N = len(x) + if N < 3: + raise ValueError("Data must be at least length 3.") + + a = zeros(N//2, dtype=np.float64) + init = 0 + + y = sort(x) + y -= x[N//2] # subtract the median (or a nearby value); see gh-15777 + + w, pw, ifault = swilk(y, a, init) + if ifault not in [0, 2]: + warnings.warn("scipy.stats.shapiro: Input data has range zero. The" + " results may not be accurate.", stacklevel=2) + if N > 5000: + warnings.warn("scipy.stats.shapiro: For N > 5000, computed p-value " + f"may not be accurate. Current N is {N}.", + stacklevel=2) + + # `w` and `pw` are always Python floats, which are double precision. + # We want to ensure that they are NumPy floats, so until dtypes are + # respected, we can explicitly convert each to float64 (faster than + # `np.array([w, pw])`). + return ShapiroResult(np.float64(w), np.float64(pw)) + + +# Values from Stephens, M A, "EDF Statistics for Goodness of Fit and +# Some Comparisons", Journal of the American Statistical +# Association, Vol. 69, Issue 347, Sept. 1974, pp 730-737 +_Avals_norm = array([0.576, 0.656, 0.787, 0.918, 1.092]) +_Avals_expon = array([0.922, 1.078, 1.341, 1.606, 1.957]) +# From Stephens, M A, "Goodness of Fit for the Extreme Value Distribution", +# Biometrika, Vol. 64, Issue 3, Dec. 1977, pp 583-588. +_Avals_gumbel = array([0.474, 0.637, 0.757, 0.877, 1.038]) +# From Stephens, M A, "Tests of Fit for the Logistic Distribution Based +# on the Empirical Distribution Function.", Biometrika, +# Vol. 66, Issue 3, Dec. 1979, pp 591-595. +_Avals_logistic = array([0.426, 0.563, 0.660, 0.769, 0.906, 1.010]) +# From Richard A. Lockhart and Michael A. Stephens "Estimation and Tests of +# Fit for the Three-Parameter Weibull Distribution" +# Journal of the Royal Statistical Society.Series B(Methodological) +# Vol. 56, No. 3 (1994), pp. 491-500, table 1. Keys are c*100 +_Avals_weibull = [[0.292, 0.395, 0.467, 0.522, 0.617, 0.711, 0.836, 0.931], + [0.295, 0.399, 0.471, 0.527, 0.623, 0.719, 0.845, 0.941], + [0.298, 0.403, 0.476, 0.534, 0.631, 0.728, 0.856, 0.954], + [0.301, 0.408, 0.483, 0.541, 0.640, 0.738, 0.869, 0.969], + [0.305, 0.414, 0.490, 0.549, 0.650, 0.751, 0.885, 0.986], + [0.309, 0.421, 0.498, 0.559, 0.662, 0.765, 0.902, 1.007], + [0.314, 0.429, 0.508, 0.570, 0.676, 0.782, 0.923, 1.030], + [0.320, 0.438, 0.519, 0.583, 0.692, 0.802, 0.947, 1.057], + [0.327, 0.448, 0.532, 0.598, 0.711, 0.824, 0.974, 1.089], + [0.334, 0.469, 0.547, 0.615, 0.732, 0.850, 1.006, 1.125], + [0.342, 0.472, 0.563, 0.636, 0.757, 0.879, 1.043, 1.167]] +_Avals_weibull = np.array(_Avals_weibull) +_cvals_weibull = np.linspace(0, 0.5, 11) +_get_As_weibull = interpolate.interp1d(_cvals_weibull, _Avals_weibull.T, + kind='linear', bounds_error=False, + fill_value=_Avals_weibull[-1]) + + +def _weibull_fit_check(params, x): + # Refine the fit returned by `weibull_min.fit` to ensure that the first + # order necessary conditions are satisfied. If not, raise an error. + # Here, use `m` for the shape parameter to be consistent with [7] + # and avoid confusion with `c` as defined in [7]. + n = len(x) + m, u, s = params + + def dnllf_dm(m, u): + # Partial w.r.t. shape w/ optimal scale. See [7] Equation 5. + xu = x-u + return (1/m - (xu**m*np.log(xu)).sum()/(xu**m).sum() + + np.log(xu).sum()/n) + + def dnllf_du(m, u): + # Partial w.r.t. loc w/ optimal scale. See [7] Equation 6. + xu = x-u + return (m-1)/m*(xu**-1).sum() - n*(xu**(m-1)).sum()/(xu**m).sum() + + def get_scale(m, u): + # Partial w.r.t. scale solved in terms of shape and location. + # See [7] Equation 7. + return ((x-u)**m/n).sum()**(1/m) + + def dnllf(params): + # Partial derivatives of the NLLF w.r.t. parameters, i.e. + # first order necessary conditions for MLE fit. + return [dnllf_dm(*params), dnllf_du(*params)] + + suggestion = ("Maximum likelihood estimation is known to be challenging " + "for the three-parameter Weibull distribution. Consider " + "performing a custom goodness-of-fit test using " + "`scipy.stats.monte_carlo_test`.") + + if np.allclose(u, np.min(x)) or m < 1: + # The critical values provided by [7] don't seem to control the + # Type I error rate in this case. Error out. + message = ("Maximum likelihood estimation has converged to " + "a solution in which the location is equal to the minimum " + "of the data, the shape parameter is less than 2, or both. " + "The table of critical values in [7] does not " + "include this case. " + suggestion) + raise ValueError(message) + + try: + # Refine the MLE / verify that first-order necessary conditions are + # satisfied. If so, the critical values provided in [7] seem reliable. + with np.errstate(over='raise', invalid='raise'): + res = optimize.root(dnllf, params[:-1]) + + message = ("Solution of MLE first-order conditions failed: " + f"{res.message}. `anderson` cannot continue. " + suggestion) + if not res.success: + raise ValueError(message) + + except (FloatingPointError, ValueError) as e: + message = ("An error occurred while fitting the Weibull distribution " + "to the data, so `anderson` cannot continue. " + suggestion) + raise ValueError(message) from e + + m, u = res.x + s = get_scale(m, u) + return m, u, s + + +AndersonResult = _make_tuple_bunch('AndersonResult', + ['statistic', 'critical_values', + 'significance_level'], ['fit_result']) + + +def anderson(x, dist='norm'): + """Anderson-Darling test for data coming from a particular distribution. + + The Anderson-Darling test tests the null hypothesis that a sample is + drawn from a population that follows a particular distribution. + For the Anderson-Darling test, the critical values depend on + which distribution is being tested against. This function works + for normal, exponential, logistic, weibull_min, or Gumbel (Extreme Value + Type I) distributions. + + Parameters + ---------- + x : array_like + Array of sample data. + dist : {'norm', 'expon', 'logistic', 'gumbel', 'gumbel_l', 'gumbel_r', 'extreme1', 'weibull_min'}, optional + The type of distribution to test against. The default is 'norm'. + The names 'extreme1', 'gumbel_l' and 'gumbel' are synonyms for the + same distribution. + + Returns + ------- + result : AndersonResult + An object with the following attributes: + + statistic : float + The Anderson-Darling test statistic. + critical_values : list + The critical values for this distribution. + significance_level : list + The significance levels for the corresponding critical values + in percents. The function returns critical values for a + differing set of significance levels depending on the + distribution that is being tested against. + fit_result : `~scipy.stats._result_classes.FitResult` + An object containing the results of fitting the distribution to + the data. + + See Also + -------- + kstest : The Kolmogorov-Smirnov test for goodness-of-fit. + + Notes + ----- + Critical values provided are for the following significance levels: + + normal/exponential + 15%, 10%, 5%, 2.5%, 1% + logistic + 25%, 10%, 5%, 2.5%, 1%, 0.5% + gumbel_l / gumbel_r + 25%, 10%, 5%, 2.5%, 1% + weibull_min + 50%, 25%, 15%, 10%, 5%, 2.5%, 1%, 0.5% + + If the returned statistic is larger than these critical values then + for the corresponding significance level, the null hypothesis that + the data come from the chosen distribution can be rejected. + The returned statistic is referred to as 'A2' in the references. + + For `weibull_min`, maximum likelihood estimation is known to be + challenging. If the test returns successfully, then the first order + conditions for a maximum likehood estimate have been verified and + the critical values correspond relatively well to the significance levels, + provided that the sample is sufficiently large (>10 observations [7]). + However, for some data - especially data with no left tail - `anderson` + is likely to result in an error message. In this case, consider + performing a custom goodness of fit test using + `scipy.stats.monte_carlo_test`. + + References + ---------- + .. [1] https://www.itl.nist.gov/div898/handbook/prc/section2/prc213.htm + .. [2] Stephens, M. A. (1974). EDF Statistics for Goodness of Fit and + Some Comparisons, Journal of the American Statistical Association, + Vol. 69, pp. 730-737. + .. [3] Stephens, M. A. (1976). Asymptotic Results for Goodness-of-Fit + Statistics with Unknown Parameters, Annals of Statistics, Vol. 4, + pp. 357-369. + .. [4] Stephens, M. A. (1977). Goodness of Fit for the Extreme Value + Distribution, Biometrika, Vol. 64, pp. 583-588. + .. [5] Stephens, M. A. (1977). Goodness of Fit with Special Reference + to Tests for Exponentiality , Technical Report No. 262, + Department of Statistics, Stanford University, Stanford, CA. + .. [6] Stephens, M. A. (1979). Tests of Fit for the Logistic Distribution + Based on the Empirical Distribution Function, Biometrika, Vol. 66, + pp. 591-595. + .. [7] Richard A. Lockhart and Michael A. Stephens "Estimation and Tests of + Fit for the Three-Parameter Weibull Distribution" + Journal of the Royal Statistical Society.Series B(Methodological) + Vol. 56, No. 3 (1994), pp. 491-500, Table 0. + + Examples + -------- + Test the null hypothesis that a random sample was drawn from a normal + distribution (with unspecified mean and standard deviation). + + >>> import numpy as np + >>> from scipy.stats import anderson + >>> rng = np.random.default_rng() + >>> data = rng.random(size=35) + >>> res = anderson(data) + >>> res.statistic + 0.8398018749744764 + >>> res.critical_values + array([0.527, 0.6 , 0.719, 0.839, 0.998]) + >>> res.significance_level + array([15. , 10. , 5. , 2.5, 1. ]) + + The value of the statistic (barely) exceeds the critical value associated + with a significance level of 2.5%, so the null hypothesis may be rejected + at a significance level of 2.5%, but not at a significance level of 1%. + + """ # numpy/numpydoc#87 # noqa: E501 + dist = dist.lower() + if dist in {'extreme1', 'gumbel'}: + dist = 'gumbel_l' + dists = {'norm', 'expon', 'gumbel_l', + 'gumbel_r', 'logistic', 'weibull_min'} + + if dist not in dists: + raise ValueError(f"Invalid distribution; dist must be in {dists}.") + y = sort(x) + xbar = np.mean(x, axis=0) + N = len(y) + if dist == 'norm': + s = np.std(x, ddof=1, axis=0) + w = (y - xbar) / s + fit_params = xbar, s + logcdf = distributions.norm.logcdf(w) + logsf = distributions.norm.logsf(w) + sig = array([15, 10, 5, 2.5, 1]) + critical = around(_Avals_norm / (1.0 + 4.0/N - 25.0/N/N), 3) + elif dist == 'expon': + w = y / xbar + fit_params = 0, xbar + logcdf = distributions.expon.logcdf(w) + logsf = distributions.expon.logsf(w) + sig = array([15, 10, 5, 2.5, 1]) + critical = around(_Avals_expon / (1.0 + 0.6/N), 3) + elif dist == 'logistic': + def rootfunc(ab, xj, N): + a, b = ab + tmp = (xj - a) / b + tmp2 = exp(tmp) + val = [np.sum(1.0/(1+tmp2), axis=0) - 0.5*N, + np.sum(tmp*(1.0-tmp2)/(1+tmp2), axis=0) + N] + return array(val) + + sol0 = array([xbar, np.std(x, ddof=1, axis=0)]) + sol = optimize.fsolve(rootfunc, sol0, args=(x, N), xtol=1e-5) + w = (y - sol[0]) / sol[1] + fit_params = sol + logcdf = distributions.logistic.logcdf(w) + logsf = distributions.logistic.logsf(w) + sig = array([25, 10, 5, 2.5, 1, 0.5]) + critical = around(_Avals_logistic / (1.0 + 0.25/N), 3) + elif dist == 'gumbel_r': + xbar, s = distributions.gumbel_r.fit(x) + w = (y - xbar) / s + fit_params = xbar, s + logcdf = distributions.gumbel_r.logcdf(w) + logsf = distributions.gumbel_r.logsf(w) + sig = array([25, 10, 5, 2.5, 1]) + critical = around(_Avals_gumbel / (1.0 + 0.2/sqrt(N)), 3) + elif dist == 'gumbel_l': + xbar, s = distributions.gumbel_l.fit(x) + w = (y - xbar) / s + fit_params = xbar, s + logcdf = distributions.gumbel_l.logcdf(w) + logsf = distributions.gumbel_l.logsf(w) + sig = array([25, 10, 5, 2.5, 1]) + critical = around(_Avals_gumbel / (1.0 + 0.2/sqrt(N)), 3) + elif dist == 'weibull_min': + message = ("Critical values of the test statistic are given for the " + "asymptotic distribution. These may not be accurate for " + "samples with fewer than 10 observations. Consider using " + "`scipy.stats.monte_carlo_test`.") + if N < 10: + warnings.warn(message, stacklevel=2) + # [7] writes our 'c' as 'm', and they write `c = 1/m`. Use their names. + m, loc, scale = distributions.weibull_min.fit(y) + m, loc, scale = _weibull_fit_check((m, loc, scale), y) + fit_params = m, loc, scale + logcdf = stats.weibull_min(*fit_params).logcdf(y) + logsf = stats.weibull_min(*fit_params).logsf(y) + c = 1 / m # m and c are as used in [7] + sig = array([0.5, 0.75, 0.85, 0.9, 0.95, 0.975, 0.99, 0.995]) + critical = _get_As_weibull(c) + # Goodness-of-fit tests should only be used to provide evidence + # _against_ the null hypothesis. Be conservative and round up. + critical = np.round(critical + 0.0005, decimals=3) + + i = arange(1, N + 1) + A2 = -N - np.sum((2*i - 1.0) / N * (logcdf + logsf[::-1]), axis=0) + + # FitResult initializer expects an optimize result, so let's work with it + message = '`anderson` successfully fit the distribution to the data.' + res = optimize.OptimizeResult(success=True, message=message) + res.x = np.array(fit_params) + fit_result = FitResult(getattr(distributions, dist), y, + discrete=False, res=res) + + return AndersonResult(A2, critical, sig, fit_result=fit_result) + + +def _anderson_ksamp_midrank(samples, Z, Zstar, k, n, N): + """Compute A2akN equation 7 of Scholz and Stephens. + + Parameters + ---------- + samples : sequence of 1-D array_like + Array of sample arrays. + Z : array_like + Sorted array of all observations. + Zstar : array_like + Sorted array of unique observations. + k : int + Number of samples. + n : array_like + Number of observations in each sample. + N : int + Total number of observations. + + Returns + ------- + A2aKN : float + The A2aKN statistics of Scholz and Stephens 1987. + + """ + A2akN = 0. + Z_ssorted_left = Z.searchsorted(Zstar, 'left') + if N == Zstar.size: + lj = 1. + else: + lj = Z.searchsorted(Zstar, 'right') - Z_ssorted_left + Bj = Z_ssorted_left + lj / 2. + for i in arange(0, k): + s = np.sort(samples[i]) + s_ssorted_right = s.searchsorted(Zstar, side='right') + Mij = s_ssorted_right.astype(float) + fij = s_ssorted_right - s.searchsorted(Zstar, 'left') + Mij -= fij / 2. + inner = lj / float(N) * (N*Mij - Bj*n[i])**2 / (Bj*(N - Bj) - N*lj/4.) + A2akN += inner.sum() / n[i] + A2akN *= (N - 1.) / N + return A2akN + + +def _anderson_ksamp_right(samples, Z, Zstar, k, n, N): + """Compute A2akN equation 6 of Scholz & Stephens. + + Parameters + ---------- + samples : sequence of 1-D array_like + Array of sample arrays. + Z : array_like + Sorted array of all observations. + Zstar : array_like + Sorted array of unique observations. + k : int + Number of samples. + n : array_like + Number of observations in each sample. + N : int + Total number of observations. + + Returns + ------- + A2KN : float + The A2KN statistics of Scholz and Stephens 1987. + + """ + A2kN = 0. + lj = Z.searchsorted(Zstar[:-1], 'right') - Z.searchsorted(Zstar[:-1], + 'left') + Bj = lj.cumsum() + for i in arange(0, k): + s = np.sort(samples[i]) + Mij = s.searchsorted(Zstar[:-1], side='right') + inner = lj / float(N) * (N * Mij - Bj * n[i])**2 / (Bj * (N - Bj)) + A2kN += inner.sum() / n[i] + return A2kN + + +Anderson_ksampResult = _make_tuple_bunch( + 'Anderson_ksampResult', + ['statistic', 'critical_values', 'pvalue'], [] +) + + +def anderson_ksamp(samples, midrank=True, *, method=None): + """The Anderson-Darling test for k-samples. + + The k-sample Anderson-Darling test is a modification of the + one-sample Anderson-Darling test. It tests the null hypothesis + that k-samples are drawn from the same population without having + to specify the distribution function of that population. The + critical values depend on the number of samples. + + Parameters + ---------- + samples : sequence of 1-D array_like + Array of sample data in arrays. + midrank : bool, optional + Type of Anderson-Darling test which is computed. Default + (True) is the midrank test applicable to continuous and + discrete populations. If False, the right side empirical + distribution is used. + method : PermutationMethod, optional + Defines the method used to compute the p-value. If `method` is an + instance of `PermutationMethod`, the p-value is computed using + `scipy.stats.permutation_test` with the provided configuration options + and other appropriate settings. Otherwise, the p-value is interpolated + from tabulated values. + + Returns + ------- + res : Anderson_ksampResult + An object containing attributes: + + statistic : float + Normalized k-sample Anderson-Darling test statistic. + critical_values : array + The critical values for significance levels 25%, 10%, 5%, 2.5%, 1%, + 0.5%, 0.1%. + pvalue : float + The approximate p-value of the test. If `method` is not + provided, the value is floored / capped at 0.1% / 25%. + + Raises + ------ + ValueError + If fewer than 2 samples are provided, a sample is empty, or no + distinct observations are in the samples. + + See Also + -------- + ks_2samp : 2 sample Kolmogorov-Smirnov test + anderson : 1 sample Anderson-Darling test + + Notes + ----- + [1]_ defines three versions of the k-sample Anderson-Darling test: + one for continuous distributions and two for discrete + distributions, in which ties between samples may occur. The + default of this routine is to compute the version based on the + midrank empirical distribution function. This test is applicable + to continuous and discrete data. If midrank is set to False, the + right side empirical distribution is used for a test for discrete + data. According to [1]_, the two discrete test statistics differ + only slightly if a few collisions due to round-off errors occur in + the test not adjusted for ties between samples. + + The critical values corresponding to the significance levels from 0.01 + to 0.25 are taken from [1]_. p-values are floored / capped + at 0.1% / 25%. Since the range of critical values might be extended in + future releases, it is recommended not to test ``p == 0.25``, but rather + ``p >= 0.25`` (analogously for the lower bound). + + .. versionadded:: 0.14.0 + + References + ---------- + .. [1] Scholz, F. W and Stephens, M. A. (1987), K-Sample + Anderson-Darling Tests, Journal of the American Statistical + Association, Vol. 82, pp. 918-924. + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> rng = np.random.default_rng() + >>> res = stats.anderson_ksamp([rng.normal(size=50), + ... rng.normal(loc=0.5, size=30)]) + >>> res.statistic, res.pvalue + (1.974403288713695, 0.04991293614572478) + >>> res.critical_values + array([0.325, 1.226, 1.961, 2.718, 3.752, 4.592, 6.546]) + + The null hypothesis that the two random samples come from the same + distribution can be rejected at the 5% level because the returned + test value is greater than the critical value for 5% (1.961) but + not at the 2.5% level. The interpolation gives an approximate + p-value of 4.99%. + + >>> samples = [rng.normal(size=50), rng.normal(size=30), + ... rng.normal(size=20)] + >>> res = stats.anderson_ksamp(samples) + >>> res.statistic, res.pvalue + (-0.29103725200789504, 0.25) + >>> res.critical_values + array([ 0.44925884, 1.3052767 , 1.9434184 , 2.57696569, 3.41634856, + 4.07210043, 5.56419101]) + + The null hypothesis cannot be rejected for three samples from an + identical distribution. The reported p-value (25%) has been capped and + may not be very accurate (since it corresponds to the value 0.449 + whereas the statistic is -0.291). + + In such cases where the p-value is capped or when sample sizes are + small, a permutation test may be more accurate. + + >>> method = stats.PermutationMethod(n_resamples=9999, random_state=rng) + >>> res = stats.anderson_ksamp(samples, method=method) + >>> res.pvalue + 0.5254 + + """ + k = len(samples) + if (k < 2): + raise ValueError("anderson_ksamp needs at least two samples") + + samples = list(map(np.asarray, samples)) + Z = np.sort(np.hstack(samples)) + N = Z.size + Zstar = np.unique(Z) + if Zstar.size < 2: + raise ValueError("anderson_ksamp needs more than one distinct " + "observation") + + n = np.array([sample.size for sample in samples]) + if np.any(n == 0): + raise ValueError("anderson_ksamp encountered sample without " + "observations") + + if midrank: + A2kN_fun = _anderson_ksamp_midrank + else: + A2kN_fun = _anderson_ksamp_right + A2kN = A2kN_fun(samples, Z, Zstar, k, n, N) + + def statistic(*samples): + return A2kN_fun(samples, Z, Zstar, k, n, N) + + if method is not None: + res = stats.permutation_test(samples, statistic, **method._asdict(), + alternative='greater') + + H = (1. / n).sum() + hs_cs = (1. / arange(N - 1, 1, -1)).cumsum() + h = hs_cs[-1] + 1 + g = (hs_cs / arange(2, N)).sum() + + a = (4*g - 6) * (k - 1) + (10 - 6*g)*H + b = (2*g - 4)*k**2 + 8*h*k + (2*g - 14*h - 4)*H - 8*h + 4*g - 6 + c = (6*h + 2*g - 2)*k**2 + (4*h - 4*g + 6)*k + (2*h - 6)*H + 4*h + d = (2*h + 6)*k**2 - 4*h*k + sigmasq = (a*N**3 + b*N**2 + c*N + d) / ((N - 1.) * (N - 2.) * (N - 3.)) + m = k - 1 + A2 = (A2kN - m) / math.sqrt(sigmasq) + + # The b_i values are the interpolation coefficients from Table 2 + # of Scholz and Stephens 1987 + b0 = np.array([0.675, 1.281, 1.645, 1.96, 2.326, 2.573, 3.085]) + b1 = np.array([-0.245, 0.25, 0.678, 1.149, 1.822, 2.364, 3.615]) + b2 = np.array([-0.105, -0.305, -0.362, -0.391, -0.396, -0.345, -0.154]) + critical = b0 + b1 / math.sqrt(m) + b2 / m + + sig = np.array([0.25, 0.1, 0.05, 0.025, 0.01, 0.005, 0.001]) + + if A2 < critical.min() and method is None: + p = sig.max() + msg = (f"p-value capped: true value larger than {p}. Consider " + "specifying `method` " + "(e.g. `method=stats.PermutationMethod()`.)") + warnings.warn(msg, stacklevel=2) + elif A2 > critical.max() and method is None: + p = sig.min() + msg = (f"p-value floored: true value smaller than {p}. Consider " + "specifying `method` " + "(e.g. `method=stats.PermutationMethod()`.)") + warnings.warn(msg, stacklevel=2) + elif method is None: + # interpolation of probit of significance level + pf = np.polyfit(critical, log(sig), 2) + p = math.exp(np.polyval(pf, A2)) + else: + p = res.pvalue if method is not None else p + + # create result object with alias for backward compatibility + res = Anderson_ksampResult(A2, critical, p) + res.significance_level = p + return res + + +AnsariResult = namedtuple('AnsariResult', ('statistic', 'pvalue')) + + +class _ABW: + """Distribution of Ansari-Bradley W-statistic under the null hypothesis.""" + # TODO: calculate exact distribution considering ties + # We could avoid summing over more than half the frequencies, + # but initially it doesn't seem worth the extra complexity + + def __init__(self): + """Minimal initializer.""" + self.m = None + self.n = None + self.astart = None + self.total = None + self.freqs = None + + def _recalc(self, n, m): + """When necessary, recalculate exact distribution.""" + if n != self.n or m != self.m: + self.n, self.m = n, m + # distribution is NOT symmetric when m + n is odd + # n is len(x), m is len(y), and ratio of scales is defined x/y + astart, a1, _ = gscale(n, m) + self.astart = astart # minimum value of statistic + # Exact distribution of test statistic under null hypothesis + # expressed as frequencies/counts/integers to maintain precision. + # Stored as floats to avoid overflow of sums. + self.freqs = a1.astype(np.float64) + self.total = self.freqs.sum() # could calculate from m and n + # probability mass is self.freqs / self.total; + + def pmf(self, k, n, m): + """Probability mass function.""" + self._recalc(n, m) + # The convention here is that PMF at k = 12.5 is the same as at k = 12, + # -> use `floor` in case of ties. + ind = np.floor(k - self.astart).astype(int) + return self.freqs[ind] / self.total + + def cdf(self, k, n, m): + """Cumulative distribution function.""" + self._recalc(n, m) + # Null distribution derived without considering ties is + # approximate. Round down to avoid Type I error. + ind = np.ceil(k - self.astart).astype(int) + return self.freqs[:ind+1].sum() / self.total + + def sf(self, k, n, m): + """Survival function.""" + self._recalc(n, m) + # Null distribution derived without considering ties is + # approximate. Round down to avoid Type I error. + ind = np.floor(k - self.astart).astype(int) + return self.freqs[ind:].sum() / self.total + + +# Maintain state for faster repeat calls to ansari w/ method='exact' +_abw_state = _ABW() + + +@_axis_nan_policy_factory(AnsariResult, n_samples=2) +def ansari(x, y, alternative='two-sided'): + """Perform the Ansari-Bradley test for equal scale parameters. + + The Ansari-Bradley test ([1]_, [2]_) is a non-parametric test + for the equality of the scale parameter of the distributions + from which two samples were drawn. The null hypothesis states that + the ratio of the scale of the distribution underlying `x` to the scale + of the distribution underlying `y` is 1. + + Parameters + ---------- + x, y : array_like + Arrays of sample data. + alternative : {'two-sided', 'less', 'greater'}, optional + Defines the alternative hypothesis. Default is 'two-sided'. + The following options are available: + + * 'two-sided': the ratio of scales is not equal to 1. + * 'less': the ratio of scales is less than 1. + * 'greater': the ratio of scales is greater than 1. + + .. versionadded:: 1.7.0 + + Returns + ------- + statistic : float + The Ansari-Bradley test statistic. + pvalue : float + The p-value of the hypothesis test. + + See Also + -------- + fligner : A non-parametric test for the equality of k variances + mood : A non-parametric test for the equality of two scale parameters + + Notes + ----- + The p-value given is exact when the sample sizes are both less than + 55 and there are no ties, otherwise a normal approximation for the + p-value is used. + + References + ---------- + .. [1] Ansari, A. R. and Bradley, R. A. (1960) Rank-sum tests for + dispersions, Annals of Mathematical Statistics, 31, 1174-1189. + .. [2] Sprent, Peter and N.C. Smeeton. Applied nonparametric + statistical methods. 3rd ed. Chapman and Hall/CRC. 2001. + Section 5.8.2. + .. [3] Nathaniel E. Helwig "Nonparametric Dispersion and Equality + Tests" at http://users.stat.umn.edu/~helwig/notes/npde-Notes.pdf + + Examples + -------- + >>> import numpy as np + >>> from scipy.stats import ansari + >>> rng = np.random.default_rng() + + For these examples, we'll create three random data sets. The first + two, with sizes 35 and 25, are drawn from a normal distribution with + mean 0 and standard deviation 2. The third data set has size 25 and + is drawn from a normal distribution with standard deviation 1.25. + + >>> x1 = rng.normal(loc=0, scale=2, size=35) + >>> x2 = rng.normal(loc=0, scale=2, size=25) + >>> x3 = rng.normal(loc=0, scale=1.25, size=25) + + First we apply `ansari` to `x1` and `x2`. These samples are drawn + from the same distribution, so we expect the Ansari-Bradley test + should not lead us to conclude that the scales of the distributions + are different. + + >>> ansari(x1, x2) + AnsariResult(statistic=541.0, pvalue=0.9762532927399098) + + With a p-value close to 1, we cannot conclude that there is a + significant difference in the scales (as expected). + + Now apply the test to `x1` and `x3`: + + >>> ansari(x1, x3) + AnsariResult(statistic=425.0, pvalue=0.0003087020407974518) + + The probability of observing such an extreme value of the statistic + under the null hypothesis of equal scales is only 0.03087%. We take this + as evidence against the null hypothesis in favor of the alternative: + the scales of the distributions from which the samples were drawn + are not equal. + + We can use the `alternative` parameter to perform a one-tailed test. + In the above example, the scale of `x1` is greater than `x3` and so + the ratio of scales of `x1` and `x3` is greater than 1. This means + that the p-value when ``alternative='greater'`` should be near 0 and + hence we should be able to reject the null hypothesis: + + >>> ansari(x1, x3, alternative='greater') + AnsariResult(statistic=425.0, pvalue=0.0001543510203987259) + + As we can see, the p-value is indeed quite low. Use of + ``alternative='less'`` should thus yield a large p-value: + + >>> ansari(x1, x3, alternative='less') + AnsariResult(statistic=425.0, pvalue=0.9998643258449039) + + """ + if alternative not in {'two-sided', 'greater', 'less'}: + raise ValueError("'alternative' must be 'two-sided'," + " 'greater', or 'less'.") + x, y = asarray(x), asarray(y) + n = len(x) + m = len(y) + if m < 1: + raise ValueError("Not enough other observations.") + if n < 1: + raise ValueError("Not enough test observations.") + + N = m + n + xy = r_[x, y] # combine + rank = _stats_py.rankdata(xy) + symrank = amin(array((rank, N - rank + 1)), 0) + AB = np.sum(symrank[:n], axis=0) + uxy = unique(xy) + repeats = (len(uxy) != len(xy)) + exact = ((m < 55) and (n < 55) and not repeats) + if repeats and (m < 55 or n < 55): + warnings.warn("Ties preclude use of exact statistic.", stacklevel=2) + if exact: + if alternative == 'two-sided': + pval = 2.0 * np.minimum(_abw_state.cdf(AB, n, m), + _abw_state.sf(AB, n, m)) + elif alternative == 'greater': + # AB statistic is _smaller_ when ratio of scales is larger, + # so this is the opposite of the usual calculation + pval = _abw_state.cdf(AB, n, m) + else: + pval = _abw_state.sf(AB, n, m) + return AnsariResult(AB, min(1.0, pval)) + + # otherwise compute normal approximation + if N % 2: # N odd + mnAB = n * (N+1.0)**2 / 4.0 / N + varAB = n * m * (N+1.0) * (3+N**2) / (48.0 * N**2) + else: + mnAB = n * (N+2.0) / 4.0 + varAB = m * n * (N+2) * (N-2.0) / 48 / (N-1.0) + if repeats: # adjust variance estimates + # compute np.sum(tj * rj**2,axis=0) + fac = np.sum(symrank**2, axis=0) + if N % 2: # N odd + varAB = m * n * (16*N*fac - (N+1)**4) / (16.0 * N**2 * (N-1)) + else: # N even + varAB = m * n * (16*fac - N*(N+2)**2) / (16.0 * N * (N-1)) + + # Small values of AB indicate larger dispersion for the x sample. + # Large values of AB indicate larger dispersion for the y sample. + # This is opposite to the way we define the ratio of scales. see [1]_. + z = (mnAB - AB) / sqrt(varAB) + pvalue = _get_pvalue(z, distributions.norm, alternative) + return AnsariResult(AB[()], pvalue[()]) + + +BartlettResult = namedtuple('BartlettResult', ('statistic', 'pvalue')) + + +@_axis_nan_policy_factory(BartlettResult, n_samples=None) +def bartlett(*samples): + r"""Perform Bartlett's test for equal variances. + + Bartlett's test tests the null hypothesis that all input samples + are from populations with equal variances. For samples + from significantly non-normal populations, Levene's test + `levene` is more robust. + + Parameters + ---------- + sample1, sample2, ... : array_like + arrays of sample data. Only 1d arrays are accepted, they may have + different lengths. + + Returns + ------- + statistic : float + The test statistic. + pvalue : float + The p-value of the test. + + See Also + -------- + fligner : A non-parametric test for the equality of k variances + levene : A robust parametric test for equality of k variances + + Notes + ----- + Conover et al. (1981) examine many of the existing parametric and + nonparametric tests by extensive simulations and they conclude that the + tests proposed by Fligner and Killeen (1976) and Levene (1960) appear to be + superior in terms of robustness of departures from normality and power + ([3]_). + + References + ---------- + .. [1] https://www.itl.nist.gov/div898/handbook/eda/section3/eda357.htm + .. [2] Snedecor, George W. and Cochran, William G. (1989), Statistical + Methods, Eighth Edition, Iowa State University Press. + .. [3] Park, C. and Lindsay, B. G. (1999). Robust Scale Estimation and + Hypothesis Testing based on Quadratic Inference Function. Technical + Report #99-03, Center for Likelihood Studies, Pennsylvania State + University. + .. [4] Bartlett, M. S. (1937). Properties of Sufficiency and Statistical + Tests. Proceedings of the Royal Society of London. Series A, + Mathematical and Physical Sciences, Vol. 160, No.901, pp. 268-282. + .. [5] C.I. BLISS (1952), The Statistics of Bioassay: With Special + Reference to the Vitamins, pp 499-503, + :doi:`10.1016/C2013-0-12584-6`. + .. [6] B. Phipson and G. K. Smyth. "Permutation P-values Should Never Be + Zero: Calculating Exact P-values When Permutations Are Randomly + Drawn." Statistical Applications in Genetics and Molecular Biology + 9.1 (2010). + .. [7] Ludbrook, J., & Dudley, H. (1998). Why permutation tests are + superior to t and F tests in biomedical research. The American + Statistician, 52(2), 127-132. + + Examples + -------- + In [5]_, the influence of vitamin C on the tooth growth of guinea pigs + was investigated. In a control study, 60 subjects were divided into + small dose, medium dose, and large dose groups that received + daily doses of 0.5, 1.0 and 2.0 mg of vitamin C, respectively. + After 42 days, the tooth growth was measured. + + The ``small_dose``, ``medium_dose``, and ``large_dose`` arrays below record + tooth growth measurements of the three groups in microns. + + >>> import numpy as np + >>> small_dose = np.array([ + ... 4.2, 11.5, 7.3, 5.8, 6.4, 10, 11.2, 11.2, 5.2, 7, + ... 15.2, 21.5, 17.6, 9.7, 14.5, 10, 8.2, 9.4, 16.5, 9.7 + ... ]) + >>> medium_dose = np.array([ + ... 16.5, 16.5, 15.2, 17.3, 22.5, 17.3, 13.6, 14.5, 18.8, 15.5, + ... 19.7, 23.3, 23.6, 26.4, 20, 25.2, 25.8, 21.2, 14.5, 27.3 + ... ]) + >>> large_dose = np.array([ + ... 23.6, 18.5, 33.9, 25.5, 26.4, 32.5, 26.7, 21.5, 23.3, 29.5, + ... 25.5, 26.4, 22.4, 24.5, 24.8, 30.9, 26.4, 27.3, 29.4, 23 + ... ]) + + The `bartlett` statistic is sensitive to differences in variances + between the samples. + + >>> from scipy import stats + >>> res = stats.bartlett(small_dose, medium_dose, large_dose) + >>> res.statistic + 0.6654670663030519 + + The value of the statistic tends to be high when there is a large + difference in variances. + + We can test for inequality of variance among the groups by comparing the + observed value of the statistic against the null distribution: the + distribution of statistic values derived under the null hypothesis that + the population variances of the three groups are equal. + + For this test, the null distribution follows the chi-square distribution + as shown below. + + >>> import matplotlib.pyplot as plt + >>> k = 3 # number of samples + >>> dist = stats.chi2(df=k-1) + >>> val = np.linspace(0, 5, 100) + >>> pdf = dist.pdf(val) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> def plot(ax): # we'll reuse this + ... ax.plot(val, pdf, color='C0') + ... ax.set_title("Bartlett Test Null Distribution") + ... ax.set_xlabel("statistic") + ... ax.set_ylabel("probability density") + ... ax.set_xlim(0, 5) + ... ax.set_ylim(0, 1) + >>> plot(ax) + >>> plt.show() + + The comparison is quantified by the p-value: the proportion of values in + the null distribution greater than or equal to the observed value of the + statistic. + + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> pvalue = dist.sf(res.statistic) + >>> annotation = (f'p-value={pvalue:.3f}\n(shaded area)') + >>> props = dict(facecolor='black', width=1, headwidth=5, headlength=8) + >>> _ = ax.annotate(annotation, (1.5, 0.22), (2.25, 0.3), arrowprops=props) + >>> i = val >= res.statistic + >>> ax.fill_between(val[i], y1=0, y2=pdf[i], color='C0') + >>> plt.show() + + >>> res.pvalue + 0.71696121509966 + + If the p-value is "small" - that is, if there is a low probability of + sampling data from distributions with identical variances that produces + such an extreme value of the statistic - this may be taken as evidence + against the null hypothesis in favor of the alternative: the variances of + the groups are not equal. Note that: + + - The inverse is not true; that is, the test is not used to provide + evidence for the null hypothesis. + - The threshold for values that will be considered "small" is a choice that + should be made before the data is analyzed [6]_ with consideration of the + risks of both false positives (incorrectly rejecting the null hypothesis) + and false negatives (failure to reject a false null hypothesis). + - Small p-values are not evidence for a *large* effect; rather, they can + only provide evidence for a "significant" effect, meaning that they are + unlikely to have occurred under the null hypothesis. + + Note that the chi-square distribution provides the null distribution + when the observations are normally distributed. For small samples + drawn from non-normal populations, it may be more appropriate to + perform a + permutation test: Under the null hypothesis that all three samples were + drawn from the same population, each of the measurements is equally likely + to have been observed in any of the three samples. Therefore, we can form + a randomized null distribution by calculating the statistic under many + randomly-generated partitionings of the observations into the three + samples. + + >>> def statistic(*samples): + ... return stats.bartlett(*samples).statistic + >>> ref = stats.permutation_test( + ... (small_dose, medium_dose, large_dose), statistic, + ... permutation_type='independent', alternative='greater' + ... ) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> bins = np.linspace(0, 5, 25) + >>> ax.hist( + ... ref.null_distribution, bins=bins, density=True, facecolor="C1" + ... ) + >>> ax.legend(['aymptotic approximation\n(many observations)', + ... 'randomized null distribution']) + >>> plot(ax) + >>> plt.show() + + >>> ref.pvalue # randomized test p-value + 0.5387 # may vary + + Note that there is significant disagreement between the p-value calculated + here and the asymptotic approximation returned by `bartlett` above. + The statistical inferences that can be drawn rigorously from a permutation + test are limited; nonetheless, they may be the preferred approach in many + circumstances [7]_. + + Following is another generic example where the null hypothesis would be + rejected. + + Test whether the lists `a`, `b` and `c` come from populations + with equal variances. + + >>> a = [8.88, 9.12, 9.04, 8.98, 9.00, 9.08, 9.01, 8.85, 9.06, 8.99] + >>> b = [8.88, 8.95, 9.29, 9.44, 9.15, 9.58, 8.36, 9.18, 8.67, 9.05] + >>> c = [8.95, 9.12, 8.95, 8.85, 9.03, 8.84, 9.07, 8.98, 8.86, 8.98] + >>> stat, p = stats.bartlett(a, b, c) + >>> p + 1.1254782518834628e-05 + + The very small p-value suggests that the populations do not have equal + variances. + + This is not surprising, given that the sample variance of `b` is much + larger than that of `a` and `c`: + + >>> [np.var(x, ddof=1) for x in [a, b, c]] + [0.007054444444444413, 0.13073888888888888, 0.008890000000000002] + + """ + k = len(samples) + if k < 2: + raise ValueError("Must enter at least two input sample vectors.") + + # Handle empty input and input that is not 1d + for sample in samples: + if np.asanyarray(sample).size == 0: + NaN = _get_nan(*samples) # get NaN of result_dtype of all samples + return BartlettResult(NaN, NaN) + + Ni = np.empty(k) + ssq = np.empty(k, 'd') + for j in range(k): + Ni[j] = len(samples[j]) + ssq[j] = np.var(samples[j], ddof=1) + Ntot = np.sum(Ni, axis=0) + spsq = np.sum((Ni - 1)*ssq, axis=0) / (1.0*(Ntot - k)) + numer = (Ntot*1.0 - k) * log(spsq) - np.sum((Ni - 1.0)*log(ssq), axis=0) + denom = 1.0 + 1.0/(3*(k - 1)) * ((np.sum(1.0/(Ni - 1.0), axis=0)) - + 1.0/(Ntot - k)) + T = numer / denom + pval = distributions.chi2.sf(T, k - 1) # 1 - cdf + + return BartlettResult(T, pval) + + +LeveneResult = namedtuple('LeveneResult', ('statistic', 'pvalue')) + + +@_axis_nan_policy_factory(LeveneResult, n_samples=None) +def levene(*samples, center='median', proportiontocut=0.05): + r"""Perform Levene test for equal variances. + + The Levene test tests the null hypothesis that all input samples + are from populations with equal variances. Levene's test is an + alternative to Bartlett's test `bartlett` in the case where + there are significant deviations from normality. + + Parameters + ---------- + sample1, sample2, ... : array_like + The sample data, possibly with different lengths. Only one-dimensional + samples are accepted. + center : {'mean', 'median', 'trimmed'}, optional + Which function of the data to use in the test. The default + is 'median'. + proportiontocut : float, optional + When `center` is 'trimmed', this gives the proportion of data points + to cut from each end. (See `scipy.stats.trim_mean`.) + Default is 0.05. + + Returns + ------- + statistic : float + The test statistic. + pvalue : float + The p-value for the test. + + See Also + -------- + fligner : A non-parametric test for the equality of k variances + bartlett : A parametric test for equality of k variances in normal samples + + Notes + ----- + Three variations of Levene's test are possible. The possibilities + and their recommended usages are: + + * 'median' : Recommended for skewed (non-normal) distributions> + * 'mean' : Recommended for symmetric, moderate-tailed distributions. + * 'trimmed' : Recommended for heavy-tailed distributions. + + The test version using the mean was proposed in the original article + of Levene ([2]_) while the median and trimmed mean have been studied by + Brown and Forsythe ([3]_), sometimes also referred to as Brown-Forsythe + test. + + References + ---------- + .. [1] https://www.itl.nist.gov/div898/handbook/eda/section3/eda35a.htm + .. [2] Levene, H. (1960). In Contributions to Probability and Statistics: + Essays in Honor of Harold Hotelling, I. Olkin et al. eds., + Stanford University Press, pp. 278-292. + .. [3] Brown, M. B. and Forsythe, A. B. (1974), Journal of the American + Statistical Association, 69, 364-367 + .. [4] C.I. BLISS (1952), The Statistics of Bioassay: With Special + Reference to the Vitamins, pp 499-503, + :doi:`10.1016/C2013-0-12584-6`. + .. [5] B. Phipson and G. K. Smyth. "Permutation P-values Should Never Be + Zero: Calculating Exact P-values When Permutations Are Randomly + Drawn." Statistical Applications in Genetics and Molecular Biology + 9.1 (2010). + .. [6] Ludbrook, J., & Dudley, H. (1998). Why permutation tests are + superior to t and F tests in biomedical research. The American + Statistician, 52(2), 127-132. + + Examples + -------- + In [4]_, the influence of vitamin C on the tooth growth of guinea pigs + was investigated. In a control study, 60 subjects were divided into + small dose, medium dose, and large dose groups that received + daily doses of 0.5, 1.0 and 2.0 mg of vitamin C, respectively. + After 42 days, the tooth growth was measured. + + The ``small_dose``, ``medium_dose``, and ``large_dose`` arrays below record + tooth growth measurements of the three groups in microns. + + >>> import numpy as np + >>> small_dose = np.array([ + ... 4.2, 11.5, 7.3, 5.8, 6.4, 10, 11.2, 11.2, 5.2, 7, + ... 15.2, 21.5, 17.6, 9.7, 14.5, 10, 8.2, 9.4, 16.5, 9.7 + ... ]) + >>> medium_dose = np.array([ + ... 16.5, 16.5, 15.2, 17.3, 22.5, 17.3, 13.6, 14.5, 18.8, 15.5, + ... 19.7, 23.3, 23.6, 26.4, 20, 25.2, 25.8, 21.2, 14.5, 27.3 + ... ]) + >>> large_dose = np.array([ + ... 23.6, 18.5, 33.9, 25.5, 26.4, 32.5, 26.7, 21.5, 23.3, 29.5, + ... 25.5, 26.4, 22.4, 24.5, 24.8, 30.9, 26.4, 27.3, 29.4, 23 + ... ]) + + The `levene` statistic is sensitive to differences in variances + between the samples. + + >>> from scipy import stats + >>> res = stats.levene(small_dose, medium_dose, large_dose) + >>> res.statistic + 0.6457341109631506 + + The value of the statistic tends to be high when there is a large + difference in variances. + + We can test for inequality of variance among the groups by comparing the + observed value of the statistic against the null distribution: the + distribution of statistic values derived under the null hypothesis that + the population variances of the three groups are equal. + + For this test, the null distribution follows the F distribution as shown + below. + + >>> import matplotlib.pyplot as plt + >>> k, n = 3, 60 # number of samples, total number of observations + >>> dist = stats.f(dfn=k-1, dfd=n-k) + >>> val = np.linspace(0, 5, 100) + >>> pdf = dist.pdf(val) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> def plot(ax): # we'll reuse this + ... ax.plot(val, pdf, color='C0') + ... ax.set_title("Levene Test Null Distribution") + ... ax.set_xlabel("statistic") + ... ax.set_ylabel("probability density") + ... ax.set_xlim(0, 5) + ... ax.set_ylim(0, 1) + >>> plot(ax) + >>> plt.show() + + The comparison is quantified by the p-value: the proportion of values in + the null distribution greater than or equal to the observed value of the + statistic. + + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> pvalue = dist.sf(res.statistic) + >>> annotation = (f'p-value={pvalue:.3f}\n(shaded area)') + >>> props = dict(facecolor='black', width=1, headwidth=5, headlength=8) + >>> _ = ax.annotate(annotation, (1.5, 0.22), (2.25, 0.3), arrowprops=props) + >>> i = val >= res.statistic + >>> ax.fill_between(val[i], y1=0, y2=pdf[i], color='C0') + >>> plt.show() + + >>> res.pvalue + 0.5280694573759905 + + If the p-value is "small" - that is, if there is a low probability of + sampling data from distributions with identical variances that produces + such an extreme value of the statistic - this may be taken as evidence + against the null hypothesis in favor of the alternative: the variances of + the groups are not equal. Note that: + + - The inverse is not true; that is, the test is not used to provide + evidence for the null hypothesis. + - The threshold for values that will be considered "small" is a choice that + should be made before the data is analyzed [5]_ with consideration of the + risks of both false positives (incorrectly rejecting the null hypothesis) + and false negatives (failure to reject a false null hypothesis). + - Small p-values are not evidence for a *large* effect; rather, they can + only provide evidence for a "significant" effect, meaning that they are + unlikely to have occurred under the null hypothesis. + + Note that the F distribution provides an asymptotic approximation of the + null distribution. + For small samples, it may be more appropriate to perform a permutation + test: Under the null hypothesis that all three samples were drawn from + the same population, each of the measurements is equally likely to have + been observed in any of the three samples. Therefore, we can form a + randomized null distribution by calculating the statistic under many + randomly-generated partitionings of the observations into the three + samples. + + >>> def statistic(*samples): + ... return stats.levene(*samples).statistic + >>> ref = stats.permutation_test( + ... (small_dose, medium_dose, large_dose), statistic, + ... permutation_type='independent', alternative='greater' + ... ) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> bins = np.linspace(0, 5, 25) + >>> ax.hist( + ... ref.null_distribution, bins=bins, density=True, facecolor="C1" + ... ) + >>> ax.legend(['aymptotic approximation\n(many observations)', + ... 'randomized null distribution']) + >>> plot(ax) + >>> plt.show() + + >>> ref.pvalue # randomized test p-value + 0.4559 # may vary + + Note that there is significant disagreement between the p-value calculated + here and the asymptotic approximation returned by `levene` above. + The statistical inferences that can be drawn rigorously from a permutation + test are limited; nonetheless, they may be the preferred approach in many + circumstances [6]_. + + Following is another generic example where the null hypothesis would be + rejected. + + Test whether the lists `a`, `b` and `c` come from populations + with equal variances. + + >>> a = [8.88, 9.12, 9.04, 8.98, 9.00, 9.08, 9.01, 8.85, 9.06, 8.99] + >>> b = [8.88, 8.95, 9.29, 9.44, 9.15, 9.58, 8.36, 9.18, 8.67, 9.05] + >>> c = [8.95, 9.12, 8.95, 8.85, 9.03, 8.84, 9.07, 8.98, 8.86, 8.98] + >>> stat, p = stats.levene(a, b, c) + >>> p + 0.002431505967249681 + + The small p-value suggests that the populations do not have equal + variances. + + This is not surprising, given that the sample variance of `b` is much + larger than that of `a` and `c`: + + >>> [np.var(x, ddof=1) for x in [a, b, c]] + [0.007054444444444413, 0.13073888888888888, 0.008890000000000002] + + """ + if center not in ['mean', 'median', 'trimmed']: + raise ValueError("center must be 'mean', 'median' or 'trimmed'.") + + k = len(samples) + if k < 2: + raise ValueError("Must enter at least two input sample vectors.") + + Ni = np.empty(k) + Yci = np.empty(k, 'd') + + if center == 'median': + + def func(x): + return np.median(x, axis=0) + + elif center == 'mean': + + def func(x): + return np.mean(x, axis=0) + + else: # center == 'trimmed' + samples = tuple(_stats_py.trimboth(np.sort(sample), proportiontocut) + for sample in samples) + + def func(x): + return np.mean(x, axis=0) + + for j in range(k): + Ni[j] = len(samples[j]) + Yci[j] = func(samples[j]) + Ntot = np.sum(Ni, axis=0) + + # compute Zij's + Zij = [None] * k + for i in range(k): + Zij[i] = abs(asarray(samples[i]) - Yci[i]) + + # compute Zbari + Zbari = np.empty(k, 'd') + Zbar = 0.0 + for i in range(k): + Zbari[i] = np.mean(Zij[i], axis=0) + Zbar += Zbari[i] * Ni[i] + + Zbar /= Ntot + numer = (Ntot - k) * np.sum(Ni * (Zbari - Zbar)**2, axis=0) + + # compute denom_variance + dvar = 0.0 + for i in range(k): + dvar += np.sum((Zij[i] - Zbari[i])**2, axis=0) + + denom = (k - 1.0) * dvar + + W = numer / denom + pval = distributions.f.sf(W, k-1, Ntot-k) # 1 - cdf + return LeveneResult(W, pval) + + +def _apply_func(x, g, func): + # g is list of indices into x + # separating x into different groups + # func should be applied over the groups + g = unique(r_[0, g, len(x)]) + output = [func(x[g[k]:g[k+1]]) for k in range(len(g) - 1)] + + return asarray(output) + + +FlignerResult = namedtuple('FlignerResult', ('statistic', 'pvalue')) + + +@_axis_nan_policy_factory(FlignerResult, n_samples=None) +def fligner(*samples, center='median', proportiontocut=0.05): + r"""Perform Fligner-Killeen test for equality of variance. + + Fligner's test tests the null hypothesis that all input samples + are from populations with equal variances. Fligner-Killeen's test is + distribution free when populations are identical [2]_. + + Parameters + ---------- + sample1, sample2, ... : array_like + Arrays of sample data. Need not be the same length. + center : {'mean', 'median', 'trimmed'}, optional + Keyword argument controlling which function of the data is used in + computing the test statistic. The default is 'median'. + proportiontocut : float, optional + When `center` is 'trimmed', this gives the proportion of data points + to cut from each end. (See `scipy.stats.trim_mean`.) + Default is 0.05. + + Returns + ------- + statistic : float + The test statistic. + pvalue : float + The p-value for the hypothesis test. + + See Also + -------- + bartlett : A parametric test for equality of k variances in normal samples + levene : A robust parametric test for equality of k variances + + Notes + ----- + As with Levene's test there are three variants of Fligner's test that + differ by the measure of central tendency used in the test. See `levene` + for more information. + + Conover et al. (1981) examine many of the existing parametric and + nonparametric tests by extensive simulations and they conclude that the + tests proposed by Fligner and Killeen (1976) and Levene (1960) appear to be + superior in terms of robustness of departures from normality and power + [3]_. + + References + ---------- + .. [1] Park, C. and Lindsay, B. G. (1999). Robust Scale Estimation and + Hypothesis Testing based on Quadratic Inference Function. Technical + Report #99-03, Center for Likelihood Studies, Pennsylvania State + University. + https://cecas.clemson.edu/~cspark/cv/paper/qif/draftqif2.pdf + .. [2] Fligner, M.A. and Killeen, T.J. (1976). Distribution-free two-sample + tests for scale. 'Journal of the American Statistical Association.' + 71(353), 210-213. + .. [3] Park, C. and Lindsay, B. G. (1999). Robust Scale Estimation and + Hypothesis Testing based on Quadratic Inference Function. Technical + Report #99-03, Center for Likelihood Studies, Pennsylvania State + University. + .. [4] Conover, W. J., Johnson, M. E. and Johnson M. M. (1981). A + comparative study of tests for homogeneity of variances, with + applications to the outer continental shelf bidding data. + Technometrics, 23(4), 351-361. + .. [5] C.I. BLISS (1952), The Statistics of Bioassay: With Special + Reference to the Vitamins, pp 499-503, + :doi:`10.1016/C2013-0-12584-6`. + .. [6] B. Phipson and G. K. Smyth. "Permutation P-values Should Never Be + Zero: Calculating Exact P-values When Permutations Are Randomly + Drawn." Statistical Applications in Genetics and Molecular Biology + 9.1 (2010). + .. [7] Ludbrook, J., & Dudley, H. (1998). Why permutation tests are + superior to t and F tests in biomedical research. The American + Statistician, 52(2), 127-132. + + Examples + -------- + In [5]_, the influence of vitamin C on the tooth growth of guinea pigs + was investigated. In a control study, 60 subjects were divided into + small dose, medium dose, and large dose groups that received + daily doses of 0.5, 1.0 and 2.0 mg of vitamin C, respectively. + After 42 days, the tooth growth was measured. + + The ``small_dose``, ``medium_dose``, and ``large_dose`` arrays below record + tooth growth measurements of the three groups in microns. + + >>> import numpy as np + >>> small_dose = np.array([ + ... 4.2, 11.5, 7.3, 5.8, 6.4, 10, 11.2, 11.2, 5.2, 7, + ... 15.2, 21.5, 17.6, 9.7, 14.5, 10, 8.2, 9.4, 16.5, 9.7 + ... ]) + >>> medium_dose = np.array([ + ... 16.5, 16.5, 15.2, 17.3, 22.5, 17.3, 13.6, 14.5, 18.8, 15.5, + ... 19.7, 23.3, 23.6, 26.4, 20, 25.2, 25.8, 21.2, 14.5, 27.3 + ... ]) + >>> large_dose = np.array([ + ... 23.6, 18.5, 33.9, 25.5, 26.4, 32.5, 26.7, 21.5, 23.3, 29.5, + ... 25.5, 26.4, 22.4, 24.5, 24.8, 30.9, 26.4, 27.3, 29.4, 23 + ... ]) + + The `fligner` statistic is sensitive to differences in variances + between the samples. + + >>> from scipy import stats + >>> res = stats.fligner(small_dose, medium_dose, large_dose) + >>> res.statistic + 1.3878943408857916 + + The value of the statistic tends to be high when there is a large + difference in variances. + + We can test for inequality of variance among the groups by comparing the + observed value of the statistic against the null distribution: the + distribution of statistic values derived under the null hypothesis that + the population variances of the three groups are equal. + + For this test, the null distribution follows the chi-square distribution + as shown below. + + >>> import matplotlib.pyplot as plt + >>> k = 3 # number of samples + >>> dist = stats.chi2(df=k-1) + >>> val = np.linspace(0, 8, 100) + >>> pdf = dist.pdf(val) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> def plot(ax): # we'll reuse this + ... ax.plot(val, pdf, color='C0') + ... ax.set_title("Fligner Test Null Distribution") + ... ax.set_xlabel("statistic") + ... ax.set_ylabel("probability density") + ... ax.set_xlim(0, 8) + ... ax.set_ylim(0, 0.5) + >>> plot(ax) + >>> plt.show() + + The comparison is quantified by the p-value: the proportion of values in + the null distribution greater than or equal to the observed value of the + statistic. + + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> pvalue = dist.sf(res.statistic) + >>> annotation = (f'p-value={pvalue:.4f}\n(shaded area)') + >>> props = dict(facecolor='black', width=1, headwidth=5, headlength=8) + >>> _ = ax.annotate(annotation, (1.5, 0.22), (2.25, 0.3), arrowprops=props) + >>> i = val >= res.statistic + >>> ax.fill_between(val[i], y1=0, y2=pdf[i], color='C0') + >>> plt.show() + + >>> res.pvalue + 0.49960016501182125 + + If the p-value is "small" - that is, if there is a low probability of + sampling data from distributions with identical variances that produces + such an extreme value of the statistic - this may be taken as evidence + against the null hypothesis in favor of the alternative: the variances of + the groups are not equal. Note that: + + - The inverse is not true; that is, the test is not used to provide + evidence for the null hypothesis. + - The threshold for values that will be considered "small" is a choice that + should be made before the data is analyzed [6]_ with consideration of the + risks of both false positives (incorrectly rejecting the null hypothesis) + and false negatives (failure to reject a false null hypothesis). + - Small p-values are not evidence for a *large* effect; rather, they can + only provide evidence for a "significant" effect, meaning that they are + unlikely to have occurred under the null hypothesis. + + Note that the chi-square distribution provides an asymptotic approximation + of the null distribution. + For small samples, it may be more appropriate to perform a + permutation test: Under the null hypothesis that all three samples were + drawn from the same population, each of the measurements is equally likely + to have been observed in any of the three samples. Therefore, we can form + a randomized null distribution by calculating the statistic under many + randomly-generated partitionings of the observations into the three + samples. + + >>> def statistic(*samples): + ... return stats.fligner(*samples).statistic + >>> ref = stats.permutation_test( + ... (small_dose, medium_dose, large_dose), statistic, + ... permutation_type='independent', alternative='greater' + ... ) + >>> fig, ax = plt.subplots(figsize=(8, 5)) + >>> plot(ax) + >>> bins = np.linspace(0, 8, 25) + >>> ax.hist( + ... ref.null_distribution, bins=bins, density=True, facecolor="C1" + ... ) + >>> ax.legend(['aymptotic approximation\n(many observations)', + ... 'randomized null distribution']) + >>> plot(ax) + >>> plt.show() + + >>> ref.pvalue # randomized test p-value + 0.4332 # may vary + + Note that there is significant disagreement between the p-value calculated + here and the asymptotic approximation returned by `fligner` above. + The statistical inferences that can be drawn rigorously from a permutation + test are limited; nonetheless, they may be the preferred approach in many + circumstances [7]_. + + Following is another generic example where the null hypothesis would be + rejected. + + Test whether the lists `a`, `b` and `c` come from populations + with equal variances. + + >>> a = [8.88, 9.12, 9.04, 8.98, 9.00, 9.08, 9.01, 8.85, 9.06, 8.99] + >>> b = [8.88, 8.95, 9.29, 9.44, 9.15, 9.58, 8.36, 9.18, 8.67, 9.05] + >>> c = [8.95, 9.12, 8.95, 8.85, 9.03, 8.84, 9.07, 8.98, 8.86, 8.98] + >>> stat, p = stats.fligner(a, b, c) + >>> p + 0.00450826080004775 + + The small p-value suggests that the populations do not have equal + variances. + + This is not surprising, given that the sample variance of `b` is much + larger than that of `a` and `c`: + + >>> [np.var(x, ddof=1) for x in [a, b, c]] + [0.007054444444444413, 0.13073888888888888, 0.008890000000000002] + + """ + if center not in ['mean', 'median', 'trimmed']: + raise ValueError("center must be 'mean', 'median' or 'trimmed'.") + + k = len(samples) + if k < 2: + raise ValueError("Must enter at least two input sample vectors.") + + # Handle empty input + for sample in samples: + if sample.size == 0: + NaN = _get_nan(*samples) + return FlignerResult(NaN, NaN) + + if center == 'median': + + def func(x): + return np.median(x, axis=0) + + elif center == 'mean': + + def func(x): + return np.mean(x, axis=0) + + else: # center == 'trimmed' + samples = tuple(_stats_py.trimboth(sample, proportiontocut) + for sample in samples) + + def func(x): + return np.mean(x, axis=0) + + Ni = asarray([len(samples[j]) for j in range(k)]) + Yci = asarray([func(samples[j]) for j in range(k)]) + Ntot = np.sum(Ni, axis=0) + # compute Zij's + Zij = [abs(asarray(samples[i]) - Yci[i]) for i in range(k)] + allZij = [] + g = [0] + for i in range(k): + allZij.extend(list(Zij[i])) + g.append(len(allZij)) + + ranks = _stats_py.rankdata(allZij) + sample = distributions.norm.ppf(ranks / (2*(Ntot + 1.0)) + 0.5) + + # compute Aibar + Aibar = _apply_func(sample, g, np.sum) / Ni + anbar = np.mean(sample, axis=0) + varsq = np.var(sample, axis=0, ddof=1) + Xsq = np.sum(Ni * (asarray(Aibar) - anbar)**2.0, axis=0) / varsq + pval = distributions.chi2.sf(Xsq, k - 1) # 1 - cdf + return FlignerResult(Xsq, pval) + + +@_axis_nan_policy_factory(lambda x1: (x1,), n_samples=4, n_outputs=1) +def _mood_inner_lc(xy, x, diffs, sorted_xy, n, m, N) -> float: + # Obtain the unique values and their frequencies from the pooled samples. + # "a_j, + b_j, = t_j, for j = 1, ... k" where `k` is the number of unique + # classes, and "[t]he number of values associated with the x's and y's in + # the jth class will be denoted by a_j, and b_j respectively." + # (Mielke, 312) + # Reuse previously computed sorted array and `diff` arrays to obtain the + # unique values and counts. Prepend `diffs` with a non-zero to indicate + # that the first element should be marked as not matching what preceded it. + diffs_prep = np.concatenate(([1], diffs)) + # Unique elements are where the was a difference between elements in the + # sorted array + uniques = sorted_xy[diffs_prep != 0] + # The count of each element is the bin size for each set of consecutive + # differences where the difference is zero. Replace nonzero differences + # with 1 and then use the cumulative sum to count the indices. + t = np.bincount(np.cumsum(np.asarray(diffs_prep != 0, dtype=int)))[1:] + k = len(uniques) + js = np.arange(1, k + 1, dtype=int) + # the `b` array mentioned in the paper is not used, outside of the + # calculation of `t`, so we do not need to calculate it separately. Here + # we calculate `a`. In plain language, `a[j]` is the number of values in + # `x` that equal `uniques[j]`. + sorted_xyx = np.sort(np.concatenate((xy, x))) + diffs = np.diff(sorted_xyx) + diffs_prep = np.concatenate(([1], diffs)) + diff_is_zero = np.asarray(diffs_prep != 0, dtype=int) + xyx_counts = np.bincount(np.cumsum(diff_is_zero))[1:] + a = xyx_counts - t + # "Define .. a_0 = b_0 = t_0 = S_0 = 0" (Mielke 312) so we shift `a` + # and `t` arrays over 1 to allow a first element of 0 to accommodate this + # indexing. + t = np.concatenate(([0], t)) + a = np.concatenate(([0], a)) + # S is built from `t`, so it does not need a preceding zero added on. + S = np.cumsum(t) + # define a copy of `S` with a prepending zero for later use to avoid + # the need for indexing. + S_i_m1 = np.concatenate(([0], S[:-1])) + + # Psi, as defined by the 6th unnumbered equation on page 313 (Mielke). + # Note that in the paper there is an error where the denominator `2` is + # squared when it should be the entire equation. + def psi(indicator): + return (indicator - (N + 1)/2)**2 + + # define summation range for use in calculation of phi, as seen in sum + # in the unnumbered equation on the bottom of page 312 (Mielke). + s_lower = S[js - 1] + 1 + s_upper = S[js] + 1 + phi_J = [np.arange(s_lower[idx], s_upper[idx]) for idx in range(k)] + + # for every range in the above array, determine the sum of psi(I) for + # every element in the range. Divide all the sums by `t`. Following the + # last unnumbered equation on page 312. + phis = [np.sum(psi(I_j)) for I_j in phi_J] / t[js] + + # `T` is equal to a[j] * phi[j], per the first unnumbered equation on + # page 312. `phis` is already in the order based on `js`, so we index + # into `a` with `js` as well. + T = sum(phis * a[js]) + + # The approximate statistic + E_0_T = n * (N * N - 1) / 12 + + varM = (m * n * (N + 1.0) * (N ** 2 - 4) / 180 - + m * n / (180 * N * (N - 1)) * np.sum( + t * (t**2 - 1) * (t**2 - 4 + (15 * (N - S - S_i_m1) ** 2)) + )) + + return ((T - E_0_T) / np.sqrt(varM),) + + +def _mood_too_small(samples, kwargs, axis=-1): + x, y = samples + n = x.shape[axis] + m = y.shape[axis] + N = m + n + return N < 3 + + +@_axis_nan_policy_factory(SignificanceResult, n_samples=2, too_small=_mood_too_small) +def mood(x, y, axis=0, alternative="two-sided"): + """Perform Mood's test for equal scale parameters. + + Mood's two-sample test for scale parameters is a non-parametric + test for the null hypothesis that two samples are drawn from the + same distribution with the same scale parameter. + + Parameters + ---------- + x, y : array_like + Arrays of sample data. + axis : int, optional + The axis along which the samples are tested. `x` and `y` can be of + different length along `axis`. + If `axis` is None, `x` and `y` are flattened and the test is done on + all values in the flattened arrays. + alternative : {'two-sided', 'less', 'greater'}, optional + Defines the alternative hypothesis. Default is 'two-sided'. + The following options are available: + + * 'two-sided': the scales of the distributions underlying `x` and `y` + are different. + * 'less': the scale of the distribution underlying `x` is less than + the scale of the distribution underlying `y`. + * 'greater': the scale of the distribution underlying `x` is greater + than the scale of the distribution underlying `y`. + + .. versionadded:: 1.7.0 + + Returns + ------- + res : SignificanceResult + An object containing attributes: + + statistic : scalar or ndarray + The z-score for the hypothesis test. For 1-D inputs a scalar is + returned. + pvalue : scalar ndarray + The p-value for the hypothesis test. + + See Also + -------- + fligner : A non-parametric test for the equality of k variances + ansari : A non-parametric test for the equality of 2 variances + bartlett : A parametric test for equality of k variances in normal samples + levene : A parametric test for equality of k variances + + Notes + ----- + The data are assumed to be drawn from probability distributions ``f(x)`` + and ``f(x/s) / s`` respectively, for some probability density function f. + The null hypothesis is that ``s == 1``. + + For multi-dimensional arrays, if the inputs are of shapes + ``(n0, n1, n2, n3)`` and ``(n0, m1, n2, n3)``, then if ``axis=1``, the + resulting z and p values will have shape ``(n0, n2, n3)``. Note that + ``n1`` and ``m1`` don't have to be equal, but the other dimensions do. + + References + ---------- + [1] Mielke, Paul W. "Note on Some Squared Rank Tests with Existing Ties." + Technometrics, vol. 9, no. 2, 1967, pp. 312-14. JSTOR, + https://doi.org/10.2307/1266427. Accessed 18 May 2022. + + Examples + -------- + >>> import numpy as np + >>> from scipy import stats + >>> rng = np.random.default_rng() + >>> x2 = rng.standard_normal((2, 45, 6, 7)) + >>> x1 = rng.standard_normal((2, 30, 6, 7)) + >>> res = stats.mood(x1, x2, axis=1) + >>> res.pvalue.shape + (2, 6, 7) + + Find the number of points where the difference in scale is not significant: + + >>> (res.pvalue > 0.1).sum() + 78 + + Perform the test with different scales: + + >>> x1 = rng.standard_normal((2, 30)) + >>> x2 = rng.standard_normal((2, 35)) * 10.0 + >>> stats.mood(x1, x2, axis=1) + SignificanceResult(statistic=array([-5.76174136, -6.12650783]), + pvalue=array([8.32505043e-09, 8.98287869e-10])) + + """ + x = np.asarray(x, dtype=float) + y = np.asarray(y, dtype=float) + + if axis < 0: + axis = x.ndim + axis + + # Determine shape of the result arrays + res_shape = tuple([x.shape[ax] for ax in range(len(x.shape)) if ax != axis]) + if not (res_shape == tuple([y.shape[ax] for ax in range(len(y.shape)) if + ax != axis])): + raise ValueError("Dimensions of x and y on all axes except `axis` " + "should match") + + n = x.shape[axis] + m = y.shape[axis] + N = m + n + if N < 3: + raise ValueError("Not enough observations.") + + xy = np.concatenate((x, y), axis=axis) + # determine if any of the samples contain ties + sorted_xy = np.sort(xy, axis=axis) + diffs = np.diff(sorted_xy, axis=axis) + if 0 in diffs: + z = np.asarray(_mood_inner_lc(xy, x, diffs, sorted_xy, n, m, N, + axis=axis)) + else: + if axis != 0: + xy = np.moveaxis(xy, axis, 0) + + xy = xy.reshape(xy.shape[0], -1) + # Generalized to the n-dimensional case by adding the axis argument, + # and using for loops, since rankdata is not vectorized. For improving + # performance consider vectorizing rankdata function. + all_ranks = np.empty_like(xy) + for j in range(xy.shape[1]): + all_ranks[:, j] = _stats_py.rankdata(xy[:, j]) + + Ri = all_ranks[:n] + M = np.sum((Ri - (N + 1.0) / 2) ** 2, axis=0) + # Approx stat. + mnM = n * (N * N - 1.0) / 12 + varM = m * n * (N + 1.0) * (N + 2) * (N - 2) / 180 + z = (M - mnM) / sqrt(varM) + pval = _get_pvalue(z, distributions.norm, alternative) + + if res_shape == (): + # Return scalars, not 0-D arrays + z = z[0] + pval = pval[0] + else: + z.shape = res_shape + pval.shape = res_shape + return SignificanceResult(z[()], pval[()]) + + +WilcoxonResult = _make_tuple_bunch('WilcoxonResult', ['statistic', 'pvalue']) + + +def wilcoxon_result_unpacker(res): + if hasattr(res, 'zstatistic'): + return res.statistic, res.pvalue, res.zstatistic + else: + return res.statistic, res.pvalue + + +def wilcoxon_result_object(statistic, pvalue, zstatistic=None): + res = WilcoxonResult(statistic, pvalue) + if zstatistic is not None: + res.zstatistic = zstatistic + return res + + +def wilcoxon_outputs(kwds): + method = kwds.get('method', 'auto') + if method == 'approx': + return 3 + return 2 + + +@_rename_parameter("mode", "method") +@_axis_nan_policy_factory( + wilcoxon_result_object, paired=True, + n_samples=lambda kwds: 2 if kwds.get('y', None) is not None else 1, + result_to_tuple=wilcoxon_result_unpacker, n_outputs=wilcoxon_outputs, +) +def wilcoxon(x, y=None, zero_method="wilcox", correction=False, + alternative="two-sided", method='auto', *, axis=0): + """Calculate the Wilcoxon signed-rank test. + + The Wilcoxon signed-rank test tests the null hypothesis that two + related paired samples come from the same distribution. In particular, + it tests whether the distribution of the differences ``x - y`` is symmetric + about zero. It is a non-parametric version of the paired T-test. + + Parameters + ---------- + x : array_like + Either the first set of measurements (in which case ``y`` is the second + set of measurements), or the differences between two sets of + measurements (in which case ``y`` is not to be specified.) Must be + one-dimensional. + y : array_like, optional + Either the second set of measurements (if ``x`` is the first set of + measurements), or not specified (if ``x`` is the differences between + two sets of measurements.) Must be one-dimensional. + + .. warning:: + When `y` is provided, `wilcoxon` calculates the test statistic + based on the ranks of the absolute values of ``d = x - y``. + Roundoff error in the subtraction can result in elements of ``d`` + being assigned different ranks even when they would be tied with + exact arithmetic. Rather than passing `x` and `y` separately, + consider computing the difference ``x - y``, rounding as needed to + ensure that only truly unique elements are numerically distinct, + and passing the result as `x`, leaving `y` at the default (None). + + zero_method : {"wilcox", "pratt", "zsplit"}, optional + There are different conventions for handling pairs of observations + with equal values ("zero-differences", or "zeros"). + + * "wilcox": Discards all zero-differences (default); see [4]_. + * "pratt": Includes zero-differences in the ranking process, + but drops the ranks of the zeros (more conservative); see [3]_. + In this case, the normal approximation is adjusted as in [5]_. + * "zsplit": Includes zero-differences in the ranking process and + splits the zero rank between positive and negative ones. + + correction : bool, optional + If True, apply continuity correction by adjusting the Wilcoxon rank + statistic by 0.5 towards the mean value when computing the + z-statistic if a normal approximation is used. Default is False. + alternative : {"two-sided", "greater", "less"}, optional + Defines the alternative hypothesis. Default is 'two-sided'. + In the following, let ``d`` represent the difference between the paired + samples: ``d = x - y`` if both ``x`` and ``y`` are provided, or + ``d = x`` otherwise. + + * 'two-sided': the distribution underlying ``d`` is not symmetric + about zero. + * 'less': the distribution underlying ``d`` is stochastically less + than a distribution symmetric about zero. + * 'greater': the distribution underlying ``d`` is stochastically + greater than a distribution symmetric about zero. + + method : {"auto", "exact", "approx"} or `PermutationMethod` instance, optional + Method to calculate the p-value, see Notes. Default is "auto". + + axis : int or None, default: 0 + If an int, the axis of the input along which to compute the statistic. + The statistic of each axis-slice (e.g. row) of the input will appear + in a corresponding element of the output. If ``None``, the input will + be raveled before computing the statistic. + + Returns + ------- + An object with the following attributes. + + statistic : array_like + If `alternative` is "two-sided", the sum of the ranks of the + differences above or below zero, whichever is smaller. + Otherwise the sum of the ranks of the differences above zero. + pvalue : array_like + The p-value for the test depending on `alternative` and `method`. + zstatistic : array_like + When ``method = 'approx'``, this is the normalized z-statistic:: + + z = (T - mn - d) / se + + where ``T`` is `statistic` as defined above, ``mn`` is the mean of the + distribution under the null hypothesis, ``d`` is a continuity + correction, and ``se`` is the standard error. + When ``method != 'approx'``, this attribute is not available. + + See Also + -------- + kruskal, mannwhitneyu + + Notes + ----- + In the following, let ``d`` represent the difference between the paired + samples: ``d = x - y`` if both ``x`` and ``y`` are provided, or ``d = x`` + otherwise. Assume that all elements of ``d`` are independent and + identically distributed observations, and all are distinct and nonzero. + + - When ``len(d)`` is sufficiently large, the null distribution of the + normalized test statistic (`zstatistic` above) is approximately normal, + and ``method = 'approx'`` can be used to compute the p-value. + + - When ``len(d)`` is small, the normal approximation may not be accurate, + and ``method='exact'`` is preferred (at the cost of additional + execution time). + + - The default, ``method='auto'``, selects between the two: when + ``len(d) <= 50`` and there are no zeros, the exact method is used; + otherwise, the approximate method is used. + + The presence of "ties" (i.e. not all elements of ``d`` are unique) or + "zeros" (i.e. elements of ``d`` are zero) changes the null distribution + of the test statistic, and ``method='exact'`` no longer calculates + the exact p-value. If ``method='approx'``, the z-statistic is adjusted + for more accurate comparison against the standard normal, but still, + for finite sample sizes, the standard normal is only an approximation of + the true null distribution of the z-statistic. For such situations, the + `method` parameter also accepts instances `PermutationMethod`. In this + case, the p-value is computed using `permutation_test` with the provided + configuration options and other appropriate settings. + + References + ---------- + .. [1] https://en.wikipedia.org/wiki/Wilcoxon_signed-rank_test + .. [2] Conover, W.J., Practical Nonparametric Statistics, 1971. + .. [3] Pratt, J.W., Remarks on Zeros and Ties in the Wilcoxon Signed + Rank Procedures, Journal of the American Statistical Association, + Vol. 54, 1959, pp. 655-667. :doi:`10.1080/01621459.1959.10501526` + .. [4] Wilcoxon, F., Individual Comparisons by Ranking Methods, + Biometrics Bulletin, Vol. 1, 1945, pp. 80-83. :doi:`10.2307/3001968` + .. [5] Cureton, E.E., The Normal Approximation to the Signed-Rank + Sampling Distribution When Zero Differences are Present, + Journal of the American Statistical Association, Vol. 62, 1967, + pp. 1068-1069. :doi:`10.1080/01621459.1967.10500917` + + Examples + -------- + In [4]_, the differences in height between cross- and self-fertilized + corn plants is given as follows: + + >>> d = [6, 8, 14, 16, 23, 24, 28, 29, 41, -48, 49, 56, 60, -67, 75] + + Cross-fertilized plants appear to be higher. To test the null + hypothesis that there is no height difference, we can apply the + two-sided test: + + >>> from scipy.stats import wilcoxon + >>> res = wilcoxon(d) + >>> res.statistic, res.pvalue + (24.0, 0.041259765625) + + Hence, we would reject the null hypothesis at a confidence level of 5%, + concluding that there is a difference in height between the groups. + To confirm that the median of the differences can be assumed to be + positive, we use: + + >>> res = wilcoxon(d, alternative='greater') + >>> res.statistic, res.pvalue + (96.0, 0.0206298828125) + + This shows that the null hypothesis that the median is negative can be + rejected at a confidence level of 5% in favor of the alternative that + the median is greater than zero. The p-values above are exact. Using the + normal approximation gives very similar values: + + >>> res = wilcoxon(d, method='approx') + >>> res.statistic, res.pvalue + (24.0, 0.04088813291185591) + + Note that the statistic changed to 96 in the one-sided case (the sum + of ranks of positive differences) whereas it is 24 in the two-sided + case (the minimum of sum of ranks above and below zero). + + In the example above, the differences in height between paired plants are + provided to `wilcoxon` directly. Alternatively, `wilcoxon` accepts two + samples of equal length, calculates the differences between paired + elements, then performs the test. Consider the samples ``x`` and ``y``: + + >>> import numpy as np + >>> x = np.array([0.5, 0.825, 0.375, 0.5]) + >>> y = np.array([0.525, 0.775, 0.325, 0.55]) + >>> res = wilcoxon(x, y, alternative='greater') + >>> res + WilcoxonResult(statistic=5.0, pvalue=0.5625) + + Note that had we calculated the differences by hand, the test would have + produced different results: + + >>> d = [-0.025, 0.05, 0.05, -0.05] + >>> ref = wilcoxon(d, alternative='greater') + >>> ref + WilcoxonResult(statistic=6.0, pvalue=0.4375) + + The substantial difference is due to roundoff error in the results of + ``x-y``: + + >>> d - (x-y) + array([2.08166817e-17, 6.93889390e-17, 1.38777878e-17, 4.16333634e-17]) + + Even though we expected all the elements of ``(x-y)[1:]`` to have the same + magnitude ``0.05``, they have slightly different magnitudes in practice, + and therefore are assigned different ranks in the test. Before performing + the test, consider calculating ``d`` and adjusting it as necessary to + ensure that theoretically identically values are not numerically distinct. + For example: + + >>> d2 = np.around(x - y, decimals=3) + >>> wilcoxon(d2, alternative='greater') + WilcoxonResult(statistic=6.0, pvalue=0.4375) + + """ + return _wilcoxon._wilcoxon_nd(x, y, zero_method, correction, alternative, + method, axis) + + +MedianTestResult = _make_tuple_bunch( + 'MedianTestResult', + ['statistic', 'pvalue', 'median', 'table'], [] +) + + +def median_test(*samples, ties='below', correction=True, lambda_=1, + nan_policy='propagate'): + """Perform a Mood's median test. + + Test that two or more samples come from populations with the same median. + + Let ``n = len(samples)`` be the number of samples. The "grand median" of + all the data is computed, and a contingency table is formed by + classifying the values in each sample as being above or below the grand + median. The contingency table, along with `correction` and `lambda_`, + are passed to `scipy.stats.chi2_contingency` to compute the test statistic + and p-value. + + Parameters + ---------- + sample1, sample2, ... : array_like + The set of samples. There must be at least two samples. + Each sample must be a one-dimensional sequence containing at least + one value. The samples are not required to have the same length. + ties : str, optional + Determines how values equal to the grand median are classified in + the contingency table. The string must be one of:: + + "below": + Values equal to the grand median are counted as "below". + "above": + Values equal to the grand median are counted as "above". + "ignore": + Values equal to the grand median are not counted. + + The default is "below". + correction : bool, optional + If True, *and* there are just two samples, apply Yates' correction + for continuity when computing the test statistic associated with + the contingency table. Default is True. + lambda_ : float or str, optional + By default, the statistic computed in this test is Pearson's + chi-squared statistic. `lambda_` allows a statistic from the + Cressie-Read power divergence family to be used instead. See + `power_divergence` for details. + Default is 1 (Pearson's chi-squared statistic). + nan_policy : {'propagate', 'raise', 'omit'}, optional + Defines how to handle when input contains nan. 'propagate' returns nan, + 'raise' throws an error, 'omit' performs the calculations ignoring nan + values. Default is 'propagate'. + + Returns + ------- + res : MedianTestResult + An object containing attributes: + + statistic : float + The test statistic. The statistic that is returned is determined + by `lambda_`. The default is Pearson's chi-squared statistic. + pvalue : float + The p-value of the test. + median : float + The grand median. + table : ndarray + The contingency table. The shape of the table is (2, n), where + n is the number of samples. The first row holds the counts of the + values above the grand median, and the second row holds the counts + of the values below the grand median. The table allows further + analysis with, for example, `scipy.stats.chi2_contingency`, or with + `scipy.stats.fisher_exact` if there are two samples, without having + to recompute the table. If ``nan_policy`` is "propagate" and there + are nans in the input, the return value for ``table`` is ``None``. + + See Also + -------- + kruskal : Compute the Kruskal-Wallis H-test for independent samples. + mannwhitneyu : Computes the Mann-Whitney rank test on samples x and y. + + Notes + ----- + .. versionadded:: 0.15.0 + + References + ---------- + .. [1] Mood, A. M., Introduction to the Theory of Statistics. McGraw-Hill + (1950), pp. 394-399. + .. [2] Zar, J. H., Biostatistical Analysis, 5th ed. Prentice Hall (2010). + See Sections 8.12 and 10.15. + + Examples + -------- + A biologist runs an experiment in which there are three groups of plants. + Group 1 has 16 plants, group 2 has 15 plants, and group 3 has 17 plants. + Each plant produces a number of seeds. The seed counts for each group + are:: + + Group 1: 10 14 14 18 20 22 24 25 31 31 32 39 43 43 48 49 + Group 2: 28 30 31 33 34 35 36 40 44 55 57 61 91 92 99 + Group 3: 0 3 9 22 23 25 25 33 34 34 40 45 46 48 62 67 84 + + The following code applies Mood's median test to these samples. + + >>> g1 = [10, 14, 14, 18, 20, 22, 24, 25, 31, 31, 32, 39, 43, 43, 48, 49] + >>> g2 = [28, 30, 31, 33, 34, 35, 36, 40, 44, 55, 57, 61, 91, 92, 99] + >>> g3 = [0, 3, 9, 22, 23, 25, 25, 33, 34, 34, 40, 45, 46, 48, 62, 67, 84] + >>> from scipy.stats import median_test + >>> res = median_test(g1, g2, g3) + + The median is + + >>> res.median + 34.0 + + and the contingency table is + + >>> res.table + array([[ 5, 10, 7], + [11, 5, 10]]) + + `p` is too large to conclude that the medians are not the same: + + >>> res.pvalue + 0.12609082774093244 + + The "G-test" can be performed by passing ``lambda_="log-likelihood"`` to + `median_test`. + + >>> res = median_test(g1, g2, g3, lambda_="log-likelihood") + >>> res.pvalue + 0.12224779737117837 + + The median occurs several times in the data, so we'll get a different + result if, for example, ``ties="above"`` is used: + + >>> res = median_test(g1, g2, g3, ties="above") + >>> res.pvalue + 0.063873276069553273 + + >>> res.table + array([[ 5, 11, 9], + [11, 4, 8]]) + + This example demonstrates that if the data set is not large and there + are values equal to the median, the p-value can be sensitive to the + choice of `ties`. + + """ + if len(samples) < 2: + raise ValueError('median_test requires two or more samples.') + + ties_options = ['below', 'above', 'ignore'] + if ties not in ties_options: + raise ValueError(f"invalid 'ties' option '{ties}'; 'ties' must be one " + f"of: {str(ties_options)[1:-1]}") + + data = [np.asarray(sample) for sample in samples] + + # Validate the sizes and shapes of the arguments. + for k, d in enumerate(data): + if d.size == 0: + raise ValueError("Sample %d is empty. All samples must " + "contain at least one value." % (k + 1)) + if d.ndim != 1: + raise ValueError("Sample %d has %d dimensions. All " + "samples must be one-dimensional sequences." % + (k + 1, d.ndim)) + + cdata = np.concatenate(data) + contains_nan, nan_policy = _contains_nan(cdata, nan_policy) + if contains_nan and nan_policy == 'propagate': + return MedianTestResult(np.nan, np.nan, np.nan, None) + + if contains_nan: + grand_median = np.median(cdata[~np.isnan(cdata)]) + else: + grand_median = np.median(cdata) + # When the minimum version of numpy supported by scipy is 1.9.0, + # the above if/else statement can be replaced by the single line: + # grand_median = np.nanmedian(cdata) + + # Create the contingency table. + table = np.zeros((2, len(data)), dtype=np.int64) + for k, sample in enumerate(data): + sample = sample[~np.isnan(sample)] + + nabove = count_nonzero(sample > grand_median) + nbelow = count_nonzero(sample < grand_median) + nequal = sample.size - (nabove + nbelow) + table[0, k] += nabove + table[1, k] += nbelow + if ties == "below": + table[1, k] += nequal + elif ties == "above": + table[0, k] += nequal + + # Check that no row or column of the table is all zero. + # Such a table can not be given to chi2_contingency, because it would have + # a zero in the table of expected frequencies. + rowsums = table.sum(axis=1) + if rowsums[0] == 0: + raise ValueError("All values are below the grand median (%r)." % + grand_median) + if rowsums[1] == 0: + raise ValueError("All values are above the grand median (%r)." % + grand_median) + if ties == "ignore": + # We already checked that each sample has at least one value, but it + # is possible that all those values equal the grand median. If `ties` + # is "ignore", that would result in a column of zeros in `table`. We + # check for that case here. + zero_cols = np.nonzero((table == 0).all(axis=0))[0] + if len(zero_cols) > 0: + msg = ("All values in sample %d are equal to the grand " + "median (%r), so they are ignored, resulting in an " + "empty sample." % (zero_cols[0] + 1, grand_median)) + raise ValueError(msg) + + stat, p, dof, expected = chi2_contingency(table, lambda_=lambda_, + correction=correction) + return MedianTestResult(stat, p, grand_median, table) + + +def _circfuncs_common(samples, high, low): + # Ensure samples are array-like and size is not zero + if samples.size == 0: + NaN = _get_nan(samples) + return NaN, NaN, NaN + + # Recast samples as radians that range between 0 and 2 pi and calculate + # the sine and cosine + sin_samp = sin((samples - low)*2.*pi / (high - low)) + cos_samp = cos((samples - low)*2.*pi / (high - low)) + + return samples, sin_samp, cos_samp + + +@_axis_nan_policy_factory( + lambda x: x, n_outputs=1, default_axis=None, + result_to_tuple=lambda x: (x,) +) +def circmean(samples, high=2*pi, low=0, axis=None, nan_policy='propagate'): + """Compute the circular mean for samples in a range. + + Parameters + ---------- + samples : array_like + Input array. + high : float or int, optional + High boundary for the sample range. Default is ``2*pi``. + low : float or int, optional + Low boundary for the sample range. Default is 0. + + Returns + ------- + circmean : float + Circular mean. + + See Also + -------- + circstd : Circular standard deviation. + circvar : Circular variance. + + Examples + -------- + For simplicity, all angles are printed out in degrees. + + >>> import numpy as np + >>> from scipy.stats import circmean + >>> import matplotlib.pyplot as plt + >>> angles = np.deg2rad(np.array([20, 30, 330])) + >>> circmean = circmean(angles) + >>> np.rad2deg(circmean) + 7.294976657784009 + + >>> mean = angles.mean() + >>> np.rad2deg(mean) + 126.66666666666666 + + Plot and compare the circular mean against the arithmetic mean. + + >>> plt.plot(np.cos(np.linspace(0, 2*np.pi, 500)), + ... np.sin(np.linspace(0, 2*np.pi, 500)), + ... c='k') + >>> plt.scatter(np.cos(angles), np.sin(angles), c='k') + >>> plt.scatter(np.cos(circmean), np.sin(circmean), c='b', + ... label='circmean') + >>> plt.scatter(np.cos(mean), np.sin(mean), c='r', label='mean') + >>> plt.legend() + >>> plt.axis('equal') + >>> plt.show() + + """ + samples, sin_samp, cos_samp = _circfuncs_common(samples, high, low) + sin_sum = sin_samp.sum(axis) + cos_sum = cos_samp.sum(axis) + res = arctan2(sin_sum, cos_sum) + + res = np.asarray(res) + res[res < 0] += 2*pi + res = res[()] + + return res*(high - low)/2.0/pi + low + + +@_axis_nan_policy_factory( + lambda x: x, n_outputs=1, default_axis=None, + result_to_tuple=lambda x: (x,) +) +def circvar(samples, high=2*pi, low=0, axis=None, nan_policy='propagate'): + """Compute the circular variance for samples assumed to be in a range. + + Parameters + ---------- + samples : array_like + Input array. + high : float or int, optional + High boundary for the sample range. Default is ``2*pi``. + low : float or int, optional + Low boundary for the sample range. Default is 0. + + Returns + ------- + circvar : float + Circular variance. + + See Also + -------- + circmean : Circular mean. + circstd : Circular standard deviation. + + Notes + ----- + This uses the following definition of circular variance: ``1-R``, where + ``R`` is the mean resultant vector. The + returned value is in the range [0, 1], 0 standing for no variance, and 1 + for a large variance. In the limit of small angles, this value is similar + to half the 'linear' variance. + + References + ---------- + .. [1] Fisher, N.I. *Statistical analysis of circular data*. Cambridge + University Press, 1993. + + Examples + -------- + >>> import numpy as np + >>> from scipy.stats import circvar + >>> import matplotlib.pyplot as plt + >>> samples_1 = np.array([0.072, -0.158, 0.077, 0.108, 0.286, + ... 0.133, -0.473, -0.001, -0.348, 0.131]) + >>> samples_2 = np.array([0.111, -0.879, 0.078, 0.733, 0.421, + ... 0.104, -0.136, -0.867, 0.012, 0.105]) + >>> circvar_1 = circvar(samples_1) + >>> circvar_2 = circvar(samples_2) + + Plot the samples. + + >>> fig, (left, right) = plt.subplots(ncols=2) + >>> for image in (left, right): + ... image.plot(np.cos(np.linspace(0, 2*np.pi, 500)), + ... np.sin(np.linspace(0, 2*np.pi, 500)), + ... c='k') + ... image.axis('equal') + ... image.axis('off') + >>> left.scatter(np.cos(samples_1), np.sin(samples_1), c='k', s=15) + >>> left.set_title(f"circular variance: {np.round(circvar_1, 2)!r}") + >>> right.scatter(np.cos(samples_2), np.sin(samples_2), c='k', s=15) + >>> right.set_title(f"circular variance: {np.round(circvar_2, 2)!r}") + >>> plt.show() + + """ + samples, sin_samp, cos_samp = _circfuncs_common(samples, high, low) + sin_mean = sin_samp.mean(axis) + cos_mean = cos_samp.mean(axis) + # hypot can go slightly above 1 due to rounding errors + with np.errstate(invalid='ignore'): + R = np.minimum(1, hypot(sin_mean, cos_mean)) + + res = 1. - R + return res + + +@_axis_nan_policy_factory( + lambda x: x, n_outputs=1, default_axis=None, + result_to_tuple=lambda x: (x,) +) +def circstd(samples, high=2*pi, low=0, axis=None, nan_policy='propagate', *, + normalize=False): + """ + Compute the circular standard deviation for samples assumed to be in the + range [low to high]. + + Parameters + ---------- + samples : array_like + Input array. + high : float or int, optional + High boundary for the sample range. Default is ``2*pi``. + low : float or int, optional + Low boundary for the sample range. Default is 0. + normalize : boolean, optional + If True, the returned value is equal to ``sqrt(-2*log(R))`` and does + not depend on the variable units. If False (default), the returned + value is scaled by ``((high-low)/(2*pi))``. + + Returns + ------- + circstd : float + Circular standard deviation. + + See Also + -------- + circmean : Circular mean. + circvar : Circular variance. + + Notes + ----- + This uses a definition of circular standard deviation from [1]_. + Essentially, the calculation is as follows. + + .. code-block:: python + + import numpy as np + C = np.cos(samples).mean() + S = np.sin(samples).mean() + R = np.sqrt(C**2 + S**2) + l = 2*np.pi / (high-low) + circstd = np.sqrt(-2*np.log(R)) / l + + In the limit of small angles, it returns a number close to the 'linear' + standard deviation. + + References + ---------- + .. [1] Mardia, K. V. (1972). 2. In *Statistics of Directional Data* + (pp. 18-24). Academic Press. :doi:`10.1016/C2013-0-07425-7`. + + Examples + -------- + >>> import numpy as np + >>> from scipy.stats import circstd + >>> import matplotlib.pyplot as plt + >>> samples_1 = np.array([0.072, -0.158, 0.077, 0.108, 0.286, + ... 0.133, -0.473, -0.001, -0.348, 0.131]) + >>> samples_2 = np.array([0.111, -0.879, 0.078, 0.733, 0.421, + ... 0.104, -0.136, -0.867, 0.012, 0.105]) + >>> circstd_1 = circstd(samples_1) + >>> circstd_2 = circstd(samples_2) + + Plot the samples. + + >>> fig, (left, right) = plt.subplots(ncols=2) + >>> for image in (left, right): + ... image.plot(np.cos(np.linspace(0, 2*np.pi, 500)), + ... np.sin(np.linspace(0, 2*np.pi, 500)), + ... c='k') + ... image.axis('equal') + ... image.axis('off') + >>> left.scatter(np.cos(samples_1), np.sin(samples_1), c='k', s=15) + >>> left.set_title(f"circular std: {np.round(circstd_1, 2)!r}") + >>> right.plot(np.cos(np.linspace(0, 2*np.pi, 500)), + ... np.sin(np.linspace(0, 2*np.pi, 500)), + ... c='k') + >>> right.scatter(np.cos(samples_2), np.sin(samples_2), c='k', s=15) + >>> right.set_title(f"circular std: {np.round(circstd_2, 2)!r}") + >>> plt.show() + + """ + samples, sin_samp, cos_samp = _circfuncs_common(samples, high, low) + sin_mean = sin_samp.mean(axis) # [1] (2.2.3) + cos_mean = cos_samp.mean(axis) # [1] (2.2.3) + # hypot can go slightly above 1 due to rounding errors + with np.errstate(invalid='ignore'): + R = np.minimum(1, hypot(sin_mean, cos_mean)) # [1] (2.2.4) + + res = sqrt(-2*log(R)) + if not normalize: + res *= (high-low)/(2.*pi) # [1] (2.3.14) w/ (2.3.7) + return res + + +class DirectionalStats: + def __init__(self, mean_direction, mean_resultant_length): + self.mean_direction = mean_direction + self.mean_resultant_length = mean_resultant_length + + def __repr__(self): + return (f"DirectionalStats(mean_direction={self.mean_direction}," + f" mean_resultant_length={self.mean_resultant_length})") + + +def directional_stats(samples, *, axis=0, normalize=True): + """ + Computes sample statistics for directional data. + + Computes the directional mean (also called the mean direction vector) and + mean resultant length of a sample of vectors. + + The directional mean is a measure of "preferred direction" of vector data. + It is analogous to the sample mean, but it is for use when the length of + the data is irrelevant (e.g. unit vectors). + + The mean resultant length is a value between 0 and 1 used to quantify the + dispersion of directional data: the smaller the mean resultant length, the + greater the dispersion. Several definitions of directional variance + involving the mean resultant length are given in [1]_ and [2]_. + + Parameters + ---------- + samples : array_like + Input array. Must be at least two-dimensional, and the last axis of the + input must correspond with the dimensionality of the vector space. + When the input is exactly two dimensional, this means that each row + of the data is a vector observation. + axis : int, default: 0 + Axis along which the directional mean is computed. + normalize: boolean, default: True + If True, normalize the input to ensure that each observation is a + unit vector. It the observations are already unit vectors, consider + setting this to False to avoid unnecessary computation. + + Returns + ------- + res : DirectionalStats + An object containing attributes: + + mean_direction : ndarray + Directional mean. + mean_resultant_length : ndarray + The mean resultant length [1]_. + + See Also + -------- + circmean: circular mean; i.e. directional mean for 2D *angles* + circvar: circular variance; i.e. directional variance for 2D *angles* + + Notes + ----- + This uses a definition of directional mean from [1]_. + Assuming the observations are unit vectors, the calculation is as follows. + + .. code-block:: python + + mean = samples.mean(axis=0) + mean_resultant_length = np.linalg.norm(mean) + mean_direction = mean / mean_resultant_length + + This definition is appropriate for *directional* data (i.e. vector data + for which the magnitude of each observation is irrelevant) but not + for *axial* data (i.e. vector data for which the magnitude and *sign* of + each observation is irrelevant). + + Several definitions of directional variance involving the mean resultant + length ``R`` have been proposed, including ``1 - R`` [1]_, ``1 - R**2`` + [2]_, and ``2 * (1 - R)`` [2]_. Rather than choosing one, this function + returns ``R`` as attribute `mean_resultant_length` so the user can compute + their preferred measure of dispersion. + + References + ---------- + .. [1] Mardia, Jupp. (2000). *Directional Statistics* + (p. 163). Wiley. + + .. [2] https://en.wikipedia.org/wiki/Directional_statistics + + Examples + -------- + >>> import numpy as np + >>> from scipy.stats import directional_stats + >>> data = np.array([[3, 4], # first observation, 2D vector space + ... [6, -8]]) # second observation + >>> dirstats = directional_stats(data) + >>> dirstats.mean_direction + array([1., 0.]) + + In contrast, the regular sample mean of the vectors would be influenced + by the magnitude of each observation. Furthermore, the result would not be + a unit vector. + + >>> data.mean(axis=0) + array([4.5, -2.]) + + An exemplary use case for `directional_stats` is to find a *meaningful* + center for a set of observations on a sphere, e.g. geographical locations. + + >>> data = np.array([[0.8660254, 0.5, 0.], + ... [0.8660254, -0.5, 0.]]) + >>> dirstats = directional_stats(data) + >>> dirstats.mean_direction + array([1., 0., 0.]) + + The regular sample mean on the other hand yields a result which does not + lie on the surface of the sphere. + + >>> data.mean(axis=0) + array([0.8660254, 0., 0.]) + + The function also returns the mean resultant length, which + can be used to calculate a directional variance. For example, using the + definition ``Var(z) = 1 - R`` from [2]_ where ``R`` is the + mean resultant length, we can calculate the directional variance of the + vectors in the above example as: + + >>> 1 - dirstats.mean_resultant_length + 0.13397459716167093 + """ + samples = np.asarray(samples) + if samples.ndim < 2: + raise ValueError("samples must at least be two-dimensional. " + f"Instead samples has shape: {samples.shape!r}") + samples = np.moveaxis(samples, axis, 0) + if normalize: + vectornorms = np.linalg.norm(samples, axis=-1, keepdims=True) + samples = samples/vectornorms + mean = np.mean(samples, axis=0) + mean_resultant_length = np.linalg.norm(mean, axis=-1, keepdims=True) + mean_direction = mean / mean_resultant_length + return DirectionalStats(mean_direction, + mean_resultant_length.squeeze(-1)[()]) + + +def false_discovery_control(ps, *, axis=0, method='bh'): + """Adjust p-values to control the false discovery rate. + + The false discovery rate (FDR) is the expected proportion of rejected null + hypotheses that are actually true. + If the null hypothesis is rejected when the *adjusted* p-value falls below + a specified level, the false discovery rate is controlled at that level. + + Parameters + ---------- + ps : 1D array_like + The p-values to adjust. Elements must be real numbers between 0 and 1. + axis : int + The axis along which to perform the adjustment. The adjustment is + performed independently along each axis-slice. If `axis` is None, `ps` + is raveled before performing the adjustment. + method : {'bh', 'by'} + The false discovery rate control procedure to apply: ``'bh'`` is for + Benjamini-Hochberg [1]_ (Eq. 1), ``'by'`` is for Benjaminini-Yekutieli + [2]_ (Theorem 1.3). The latter is more conservative, but it is + guaranteed to control the FDR even when the p-values are not from + independent tests. + + Returns + ------- + ps_adusted : array_like + The adjusted p-values. If the null hypothesis is rejected where these + fall below a specified level, the false discovery rate is controlled + at that level. + + See Also + -------- + combine_pvalues + statsmodels.stats.multitest.multipletests + + Notes + ----- + In multiple hypothesis testing, false discovery control procedures tend to + offer higher power than familywise error rate control procedures (e.g. + Bonferroni correction [1]_). + + If the p-values correspond with independent tests (or tests with + "positive regression dependencies" [2]_), rejecting null hypotheses + corresponding with Benjamini-Hochberg-adjusted p-values below :math:`q` + controls the false discovery rate at a level less than or equal to + :math:`q m_0 / m`, where :math:`m_0` is the number of true null hypotheses + and :math:`m` is the total number of null hypotheses tested. The same is + true even for dependent tests when the p-values are adjusted accorded to + the more conservative Benjaminini-Yekutieli procedure. + + The adjusted p-values produced by this function are comparable to those + produced by the R function ``p.adjust`` and the statsmodels function + `statsmodels.stats.multitest.multipletests`. Please consider the latter + for more advanced methods of multiple comparison correction. + + References + ---------- + .. [1] Benjamini, Yoav, and Yosef Hochberg. "Controlling the false + discovery rate: a practical and powerful approach to multiple + testing." Journal of the Royal statistical society: series B + (Methodological) 57.1 (1995): 289-300. + + .. [2] Benjamini, Yoav, and Daniel Yekutieli. "The control of the false + discovery rate in multiple testing under dependency." Annals of + statistics (2001): 1165-1188. + + .. [3] TileStats. FDR - Benjamini-Hochberg explained - Youtube. + https://www.youtube.com/watch?v=rZKa4tW2NKs. + + .. [4] Neuhaus, Karl-Ludwig, et al. "Improved thrombolysis in acute + myocardial infarction with front-loaded administration of alteplase: + results of the rt-PA-APSAC patency study (TAPS)." Journal of the + American College of Cardiology 19.5 (1992): 885-891. + + Examples + -------- + We follow the example from [1]_. + + Thrombolysis with recombinant tissue-type plasminogen activator (rt-PA) + and anisoylated plasminogen streptokinase activator (APSAC) in + myocardial infarction has been proved to reduce mortality. [4]_ + investigated the effects of a new front-loaded administration of rt-PA + versus those obtained with a standard regimen of APSAC, in a randomized + multicentre trial in 421 patients with acute myocardial infarction. + + There were four families of hypotheses tested in the study, the last of + which was "cardiac and other events after the start of thrombolitic + treatment". FDR control may be desired in this family of hypotheses + because it would not be appropriate to conclude that the front-loaded + treatment is better if it is merely equivalent to the previous treatment. + + The p-values corresponding with the 15 hypotheses in this family were + + >>> ps = [0.0001, 0.0004, 0.0019, 0.0095, 0.0201, 0.0278, 0.0298, 0.0344, + ... 0.0459, 0.3240, 0.4262, 0.5719, 0.6528, 0.7590, 1.000] + + If the chosen significance level is 0.05, we may be tempted to reject the + null hypotheses for the tests corresponding with the first nine p-values, + as the first nine p-values fall below the chosen significance level. + However, this would ignore the problem of "multiplicity": if we fail to + correct for the fact that multiple comparisons are being performed, we + are more likely to incorrectly reject true null hypotheses. + + One approach to the multiplicity problem is to control the family-wise + error rate (FWER), that is, the rate at which the null hypothesis is + rejected when it is actually true. A common procedure of this kind is the + Bonferroni correction [1]_. We begin by multiplying the p-values by the + number of hypotheses tested. + + >>> import numpy as np + >>> np.array(ps) * len(ps) + array([1.5000e-03, 6.0000e-03, 2.8500e-02, 1.4250e-01, 3.0150e-01, + 4.1700e-01, 4.4700e-01, 5.1600e-01, 6.8850e-01, 4.8600e+00, + 6.3930e+00, 8.5785e+00, 9.7920e+00, 1.1385e+01, 1.5000e+01]) + + To control the FWER at 5%, we reject only the hypotheses corresponding + with adjusted p-values less than 0.05. In this case, only the hypotheses + corresponding with the first three p-values can be rejected. According to + [1]_, these three hypotheses concerned "allergic reaction" and "two + different aspects of bleeding." + + An alternative approach is to control the false discovery rate: the + expected fraction of rejected null hypotheses that are actually true. The + advantage of this approach is that it typically affords greater power: an + increased rate of rejecting the null hypothesis when it is indeed false. To + control the false discovery rate at 5%, we apply the Benjamini-Hochberg + p-value adjustment. + + >>> from scipy import stats + >>> stats.false_discovery_control(ps) + array([0.0015 , 0.003 , 0.0095 , 0.035625 , 0.0603 , + 0.06385714, 0.06385714, 0.0645 , 0.0765 , 0.486 , + 0.58118182, 0.714875 , 0.75323077, 0.81321429, 1. ]) + + Now, the first *four* adjusted p-values fall below 0.05, so we would reject + the null hypotheses corresponding with these *four* p-values. Rejection + of the fourth null hypothesis was particularly important to the original + study as it led to the conclusion that the new treatment had a + "substantially lower in-hospital mortality rate." + + """ + # Input Validation and Special Cases + ps = np.asarray(ps) + + ps_in_range = (np.issubdtype(ps.dtype, np.number) + and np.all(ps == np.clip(ps, 0, 1))) + if not ps_in_range: + raise ValueError("`ps` must include only numbers between 0 and 1.") + + methods = {'bh', 'by'} + if method.lower() not in methods: + raise ValueError(f"Unrecognized `method` '{method}'." + f"Method must be one of {methods}.") + method = method.lower() + + if axis is None: + axis = 0 + ps = ps.ravel() + + axis = np.asarray(axis)[()] + if not np.issubdtype(axis.dtype, np.integer) or axis.size != 1: + raise ValueError("`axis` must be an integer or `None`") + + if ps.size <= 1 or ps.shape[axis] <= 1: + return ps[()] + + ps = np.moveaxis(ps, axis, -1) + m = ps.shape[-1] + + # Main Algorithm + # Equivalent to the ideas of [1] and [2], except that this adjusts the + # p-values as described in [3]. The results are similar to those produced + # by R's p.adjust. + + # "Let [ps] be the ordered observed p-values..." + order = np.argsort(ps, axis=-1) + ps = np.take_along_axis(ps, order, axis=-1) # this copies ps + + # Equation 1 of [1] rearranged to reject when p is less than specified q + i = np.arange(1, m+1) + ps *= m / i + + # Theorem 1.3 of [2] + if method == 'by': + ps *= np.sum(1 / i) + + # accounts for rejecting all null hypotheses i for i < k, where k is + # defined in Eq. 1 of either [1] or [2]. See [3]. Starting with the index j + # of the second to last element, we replace element j with element j+1 if + # the latter is smaller. + np.minimum.accumulate(ps[..., ::-1], out=ps[..., ::-1], axis=-1) + + # Restore original order of axes and data + np.put_along_axis(ps, order, values=ps.copy(), axis=-1) + ps = np.moveaxis(ps, -1, axis) + + return np.clip(ps, 0, 1)