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Open AccessArticle

Computation of Probability Associated with Anderson–Darling Statistic

1
Department of Physics and Chemistry, Technical University of Cluj-Napoca, Muncii Blvd. No. 103-105, Cluj-Napoca 400641, Romania
2
Doctoral Studies, Babeş-Bolyai University, Mihail Kogălniceanu Str., No. 1, Cluj-Napoca 400028, Romania
3
Department of Medical Informatics and Biostatistics, Iuliu Haţieganu University of Medicine and Pharmacy, Louis Pasteur Str., No. 6, Cluj-Napoca 400349, Romania
*
Author to whom correspondence should be addressed.
Mathematics 2018, 6(6), 88; https://doi.org/10.3390/math6060088
Received: 14 April 2018 / Revised: 21 May 2018 / Accepted: 23 May 2018 / Published: 25 May 2018
(This article belongs to the Special Issue Applied and Computational Statistics)

Abstract

The correct application of a statistical test is directly connected with information related to the distribution of data. Anderson–Darling is one alternative used to test if the distribution of experimental data follows a theoretical distribution. The conclusion of the Anderson–Darling test is usually drawn by comparing the obtained statistic with the available critical value, which did not give any weight to the same size. This study aimed to provide a formula for calculation of p-value associated with the Anderson–Darling statistic considering the size of the sample. A Monte Carlo simulation study was conducted for sample sizes starting from 2 to 61, and based on the obtained results, a formula able to give reliable probabilities associated to the Anderson–Darling statistic is reported.
Keywords: Anderson–Darling test (AD); probability; Monte Carlo simulation Anderson–Darling test (AD); probability; Monte Carlo simulation

1. Introduction

Application of any statistical test is made under certain assumptions, and violation of these assumptions could lead to misleading interpretations and unreliable results [1,2]. One main assumption that several statistical tests have is related with the distribution of experimental or observed data (H0 (null hypothesis): The data follow the specified distribution vs. H1 (alternative hypothesis): The data do not follow the specified distribution). Different tests, generally called “goodness-of-fit”, are used to assess whether a sample of observations can be considered as a sample from a given distribution. The most frequently used goodness-of-fit tests are Kolmogorov–Smirnov [3,4], Anderson–Darling [5,6], Pearson’s chi-square [7], Cramér–von Mises [8,9], Shapiro–Wilk [10], Jarque–Bera [11,12,13], D’Agostino–Pearson [14], and Lilliefors [15,16]. The goodness-of-fit tests use different procedures (see Table 1). Alongside the well-known goodness-of-fit test, other methods based for example on entropy estimator [17,18,19], jackknife empirical likelihood [20], on the prediction of residuals [21], or for testing multilevel survival data [22] or multilevel models with binary outcomes [23] have been reported in the scientific literature.
Tests used to assess the distribution of a dataset received attention from many researchers (for testing normal or other distributions) [24,25,26,27]. The normal distribution is of higher importance, since the resulting information will lead the statistical analysis on the pathway of parametric or non-parametric tests [28,29,30,31,32,33]. Different normality tests are implemented on various statistical packages (e.g., Minitab—http://www.minitab.com/en-us/; EasyFit—http://www.mathwave.com/easyfit-distribution-fitting.html; Develve—http://develve.net/; r(“nortest” nortest)—https://cran.r-project.org/web/packages/nortest/nortest.pdf; etc.).
Several studies aimed to compare the performances of goodness-of-fit tests. In a Monte Carlo simulation study conducted on the normal distribution, Kolmogorov–Smirnov test has been identified as the least powerful test, while opposite Shapiro–Wilks test was identified as the most powerful test [34]. Furthermore, Anderson–Darling test was found to be the best option among five normality tests whenever t-statistics were used [35]. More weight to the tails are given by the Anderson–Darling test compared to Kolmogorov–Smirnov test [36]. The comparisons between different goodness-of-fit tests is frequently conducted by comparing their power [37,38], using or not confidence intervals [39], distribution of p-values [40], or ROC (receiver operating characteristic) analysis [32].
The interpretation of the Anderson–Darling test is frequently made by comparing the AD statistic with the critical value for a particular significance level (e.g., 20%, 10%, 5%, 2.5%, or 1%) even if it is known that the critical values depend on the sample size [41,42]. The main problem with this approach is that the critical values are available just for several distributions (e.g., normal and Weibull distribution in Table 2 [43], generalized extreme value and generalized logistic [44], etc.) but could be obtained in Monte Carlo simulations [45]. The primary advantage of the Anderson–Darling test is its applicability to test the departure of the experimental data from different theoretical distributions, which is the reason why we decided to identify the method able to calculate its associated p-value as a function also of the sample size.
D’Augostino and Stephens provided different formulas for calculation of p-values associated to the Anderson–Darling statistic (AD), along with a correction for small sample size (AD*) [37]. Their equations are independent of the tested theoretical distribution and highlight the importance of the sample size (Table 3).
Several Excel implementations of Anderson–Darling statistic are freely available to assist the researcher in testing if data follow, or do not follow, the normal distribution [46,47,48]. Since almost all distributions are dependent by at least two parameters, it is not expected that one goodness-of-fit test will provide sufficient information regarding the risk of error, because using only one method (one test) gives the expression of only one constraint between parameters. In this regard, the example provided in [49] is illustrative, and shows how the presence of a single outlier induces complete disarray between statistics, and even its removal does not bring the same risk of error as a result of applying different goodness-of-fit tests. Given this fact, calculation of the combined probability of independent (e.g., independent of the tested distribution) goodness-of-fit tests [50,51] is justified.
Good statistical practice guidelines request reporting the p-value associated with the statistics of a test. The sample size influences the p-value of statistics, so its reporting is mandatory to assure a proper interpretation of the statistical results. Our study aimed to identify, assess, and implement an explicit function of the p-value associated with the Anderson–Darling statistic able to take into consideration both the value of the statistic and the sample size.

2. Materials and Methods

2.1. Anderson–Darling Order Statistic

For a sample Y = (y1, y2, …, yn), the data are sorted in ascending order (let X = Sort(Y), and then X = (x1, x2, …, xn) with xi ≤ xi+1 for 0 < i < n, and xi = yσ(i), where σ is a permutation of {1, 2, …, n} which makes the X series sorted). Let the CDF be the associated cumulative distribution function and InvCDF the inverse of this function for any PDF (probability density function). The series P = (p1, p2, …, pn) defined by pi = InvCDF(xi) (or Q = (q1, q2, …, qn) defined by qi = InvCDF(yi), where the P is the unsorted array, and Q is the sorted array) are samples drawn from a uniform distribution only if Y (and X) are samples from the distribution with PDF.
At this point, the order statistics are used to test the uniformity of P (or for Q), and for this reason, the values of X are ordered (in Y). On the ordered probabilities (on P), several statistics can be computed, and Anderson–Darling (AD) is one of them:
A D = A D ( P ,   n ) = n i = 1 n ( 2 i 1 ) ln ( p i ( 1 p n i + 1 ) ) n .
The associated AD statistic for a “perfect” uniform distribution can be computed after splitting the [0, 1] interval into n equidistant intervals (i/n, with 0 ≤ in being their boundaries) and using the middles of those intervals ri = (2i − 1)/2n:
A D min ( n ) = A D ( R , n ) = n + 4 H 1 ( R , n ) .
where H1 is the Shannon entropy for R in nats (the units of information or entropy) (H1(R,n) = − Σri∙ln(ri)).
Equation (2) gives the smallest possible value for AD. The value of the AD increases with the increase of the departure between the perfect uniform distribution and the observed distribution (P).

2.2. Monte Carlo Experiment for Anderson–Darling Statistic

The probability associated with a particular value of the AD statistic can be obtained using a Monte Carlo experiment. The AD statistics are calculated for a large enough number of samples (let be m the number of samples), the values are sorted, and then the relative position of the observed value of the AD in the series of Monte Carlo-calculated values gives the probability associated with the statistic of the AD test.
It should be noted that the equation linking the statistic and the probability also contains the size of the sample, and therefore, the probability associated with the AD value is dependent on n.
Taking into account all the knowledge gains until this point, it is relatively simple to do a Monte Carlo experiment for any order statistic. The only remaining problem is how to draw a sample from a uniform distribution in such way as to not affect the outcome. One alternative is to use a good random generator, such as Mersenne Twister [53], and this method was used to generate our samples as an alternative to the stratified random approach.

2.3. Stratified Random Strategy

Let us assume that three numbers (t1, t2, t3) are extracted from a [0, 1) interval using Mersenne Twister method. Each of those numbers can be <0.5 or ≥0.5, providing 23 possible cases (Table 4).
It is not a good idea to use the design presented in Table 4 in its crude form, since it is transformed to a problem with an exponential (2n) complexity. The trick is to observe the pattern in Table 4. In fact, for (n + 1) cases, with different frequencies of occurrence following the model, the results are given in Table 5.
The complexity of the problem of enumerating all the cases stays with the design presented in Table 5 at the same order of magnitude with n (we need to list only n + 1 cases instead of 2n).
The frequencies listed in Table 5 are combinations of n objects taken by two (intervals), so instead of enumerating all 2n cases, it is enough to record only n + 1 cases weighted with their relative occurrence.
The effect of the pseudo-random generator is significantly decreased (the decrease is a precise order of magnitude of the binary representation, one unit in log2 transformation: 1 = log22, for the (0, 0.5) and (0.5, 1) split) by doing a stratified random sample.
The extractions of a number from (0, 0.5) and from (0.5, 1) were furthermore made in our experiment with Mersenne Twister random (if x = Random() with 0 ≤ x < 1 then 0 ≤ x/2 < 1 and 0.5 ≤ 0.5 + x/2 < 1). Table 5 provides all the information we need to do the design. For any n, for k from 0 to n, exactly k numbers are generated as Random()/2, and sorted. Furthermore, exactly nk numbers are generated as 0.5 + Random()/2, and the frequency associated with this pattern is n!/(k!∙(nk)!).
The combinations can also be calculated iteratively: cnk(n,0) = 1, and cnk(n,k) = cnk(n,(k − 1))∙(nk + 1)/k for successive 1 ≤ kn.

2.4. Model for Anderson–Darling Statistic

Performing the Monte Carlo (MC) experiment (generates, analyzes, and provides the outcome) each time when a probability associated with the AD statistic is needed is resource-consuming and not effective. For example, if we generate for a certain sample size (n) a large number of samples m = 1.28 × 1010, then the needed storage space is 51.2 Gb for each n. Given 1 Tb of storage capacity, it can store only 20 iterations of n, as in the series of the AD(n). However, this is not needed, since it is possible to generate and store the results of the Monte Carlo analysis, but a proper model is required.
It is not necessary to have a model for any probability, since the standard thresholds for rejecting an agreement are commonly set to α = 0.2, 0.1, 0.05, 0.02, 0.01 (α = 1 − p). A reliable result could be considered the model for the AD when p ≥ 0.5. Therefore, the AD (as AD = AD(n,p)) for 501 value of the p from 0.500 to 0.001, and for n from 2 to 61 were extracted, tabulated, and used to develop the model.
A search for a dependency of AD = AD(p) (or p = p(AD)) for a particular n may not reveal any pattern. However, if the value of the statistic is exponentiated (see the ln function in the AD formula), values for the model start to appear (see Figure 1a) after a proper transformation of p. On the other hand, for a given n, an inconvenience for the AD(p) (or for its inverse, p = p(AD)) is to have on the plot, a non-uniform repartition of the points—for instance, precisely two points for 5 ≤ AD < 6 and 144 points for AD < 1. As a consequence, any method trying to find the best fit based on this raw data will fail because it will give too much weight on the lower part with a much higher concentration of the points. The problem is the same for exp(AD) replacing AD (Figure 1b) but is no more the case for 1/(1 − p) as a function of exp(AD) (Figure 1c), since the dependence begins to look like a linear one. Figure 1b suggests that a logarithm on both axes will reduce the difference in the concentration of points in the intervals (Figure 1d), but at this point, is not necessary to apply it, since the last spots in Figure 1c may act as “outliers” trailing the slope. A good fit in the rarefied region of high p (and low α) is desired. It is not so important if we will have a 1% error at p = 50%, but is essential not to have a 1% error at p = 99% (the error will be higher than the estimated probability, α = 1 − p. Therefore, in this case (Figure 1c), big numbers (e.g., ~200, 400) will have high values of residuals, and will trail the model to fit better in the rarefied region.
A simple linear regression y ~ ŷ = a∙x + b for x ← eAD and y ← α − 1 = 1/(1 − p) will do most of the job for providing the values of α associated with the values of the AD. Since the dependence is almost linear, polynomial or rational functions will perform worse, as proven in the tests. A better alternative is to feed the model with fractional powers of x. By doing this, the bigger numbers will not be disfavored (square root of 100 is 10, which is ten times lower than 100, while square root of 1 is 1; thus, the weight of the linear component is less affected for bigger numbers). On the other hand, looking to the AD definition, the probability is raised at a variable power, and therefore, to turn back to it, in the conventional sense of operation, is to do root. Our proposed model is given in Equation (3):
y ^ = a 0 + a 1 x 1 / 4 + a 2 x 2 / 4 + a 3 x 3 / 4 + a 4 x
The statistics associated with the proposed model for data presented in Figure 1 are given in Table 6.
The analysis of the results presented in Table 6 showed that all coefficients are statistically significant, and their significance increases from the coefficient of AD1/4 to the coefficient of the AD. Furthermore, the residuals of the regression are with ten orders of magnitude less than the total residuals (F value = 3.4 × 1010). The adjusted determination coefficient has eight consecutive nines.
The model is not finished yet, because we need a model that also embeds the sample size (n). Inverse powers of n are the best alternatives as already suggested in the literature [43]. Therefore, for each coefficient (from a0 to a4), a function penalizing the small samples was used similarly:
a ^ i = b 0 , i + b 1 , i n 1 + b 2 , i n 2 + b 3 , i n 3 + b 4 , i n 4 .
With these replacements, the whole model providing the probability as a function of AD statistic and n is given by Equation (5):
y ^ = i = 0 4 j = 0 4 b i , j x i / 4 n j ,
where ŷ = 1/(1 − p), bi,j = coefficients, x = eAD, n = sample size.

3. Simulation Results

Twenty-five coefficients were calculated for Equation (5) from 60 values associated to sample sizes from 2 to 61, based on 500 values of p (0.500 ≤ p ≤ 0.999) and with a step of 0.001. The values of the obtained coefficients along with the related Student t-statistic are given in Table 7.

3.1. Stratified vs. Random

The same experiment was conducted with both simple and random stratified Mersenne Twister method [53] to assess the magnitude of the increases in the resolution of the AD statistic. The differences between the two scenarios were calculated and plotted in Figure 2.

3.2. Analysis of Residuals

The residuals, defined as the difference between the probability obtained by Monte Carlo simulation and the value estimated by the proposed model, without and with transformation (ln and respectively log), were analyzed. For each probability (p ranging from 0.500 to 0.999 with a step of 0.001; 500 values) associated with the statistic (AD) based on the MC simulation for n ranging from 2 to 61 (60 values), 30,000 distinct pairs (p, n, AD) were collected and investigated. The descriptive statistics of residuals are presented in Table 8.
The most frequent value of residuals (~99%) equals with 0.000007 when no transformed data are investigated (Figure 3, left-hand graph). The right-hand chart in Figure 3 depicted the distribution of the same data, but expressed in logarithmical scale, showing a better agreement with normal distribution for the transformed residuals.
A sample of p ranging from 0.500 to 0.995 with a step of 0.005 (100 values), and for n in the same range (from 2 to 61; 60 values) was extracted from the whole pool of data, and a 3D mesh with 6000 grid points was constructed. Figure 4 represents the differences log10(p p ^ ) ( p ^ is calculated with Equation (5)) and the values of the bi,j coefficients given in Table 4. For convenience, the equation for p ^ and (α ≡ 1 × p) are
p ^ = 1 ( i = 0 4 j = 0 4 b i , j x i / 4 n j ) 1
a ^ = ( i = 0 4 j = 0 4 b i , j x i / 4 n j ) 1 .
Figure 4 reveals that the calculated Equation (5) and the expected values (from MC simulation for AD = AD(p,n)) differ less than 1‰ (−3 on the top of the Z axis). Even more than that, with departure from n = 2, and from p = 0.500 to n = 61, or to p = 0.999, the difference dramatically decreases to 10−6 (visible on the Z-axis as −6 moving from n = 2 to n = 61), to 10−9 (visible on the plot visible on X-axis as −9 moving from p = 0.500 to p = 0.995), and even to 10−15 (visible on the plot on Z-axis as −15 moving on both from p = 0.500 to p = 0.995 and from n = 2 to n = 61). This behavior shows that the model was designed in a way in which the estimation error (p p ^ ) would be minimal for small α (α close to 0; p close to 1). A regular half-circle shape pattern, depicted in Figure 4, suggests that an even more precise method than the one archived by the proposed model must be done with periodic functions.
Figure 5 illustrates, more obviously, this pattern with the peak at n = 2 and p = 0.500.
Median of residuals expressed in logarithmic scale indicate that half of the points have exactly seven digits (e.g., 0.98900000 vs. 0.98900004). The cumulative frequencies for the residuals represented in logarithmic scale also show that 75% have exactly six digits, while over 99% have exactly five digits. The agreement between the observed Monte Carlo and the regression model is excellent (r2(n = 30,000) = 0.99999) with a minimum value for the sum of squares of residuals (0.002485). These results sustain the validity of the proposed model.

4. Case Study

Twenty sets of experimental data (Table 9) were used to test the hypothesis of the normal distribution:
H1:
The distribution of experimental data is not significantly different from the theoretical normal distribution.
H2:
The distribution of experimental data is not significantly different from the theoretical normal distribution.
Experimental data were analyzed with EasyFit Professional (v. 5.2) [72], and the retrieved AD statistic, along with the conclusion of the test (Reject H0?) at a significance level of 5% were recorded. The AD statistic and the sample size for each dataset were used to retrieve the p-value calculated with our method. As a control method, the formulas presented in Table 3 [43], implemented in an Excel file (SPC for Excel) [47], were used. The obtained results are presented in Table 10.
A perfect concordance was observed in regard to the statistical conclusion regarding the normal distribution, when our method was compared to the judgment retrieved by EasyFit. The concordance of the results between SPC and EasyFit, respectively, with the proposed method, was 60%, with discordant results for both small (e.g., n = 24, set 1) samples as well as high (e.g., n = 70, set 9) sample sizes. Normal probability plots (P–P) and the quantile–quantile plots (Q–Q) of these sets show slight, but not significant deviations from the expected normal distribution (Figure 6).
Without any exceptions, the p-values calculated by our implemented method had higher values compared to the p-values achieved by SPC for Excel. The most substantial difference is observed for the largest dataset (set 8), while the smallest difference is noted for the set with 45 experimental data values (set 15). The lowest p-value was obtained by the implemented method for set 3 (see Table 10); the SPC for Excel retrieves, for this dataset, a value of 0.0000. The next smallest p-value was observed for set 8. For both these sets, an agreement related to the statistical decision was found (see Table 10).
Our team has previously investigated the effect of sample size on the probability of Anderson–Darling test, and the results are published online at http://l.academicdirect.org/Statistics/tests/AD/. The method proposed in this manuscript, as compared to the previous one, assures a higher resolution expressed by the lower unexplained variance between the AD and the model using a formula with a smaller number of coefficients. Furthermore, the unexplained variance of the method present in this manuscript has much less weight for big “p-values”, and much higher weight for small “p-values”, which means that is more appropriate to be used for low (e.g., p ~10−5) and very low (p ~10−10) probabilities.
Further research could be done in both the extension of the proposed method and the evaluation of its performances. The performances of the reported method could be evaluated for the whole range of sample sizes if proper computational resources exist. Furthermore, the performance of the implementation could be assessed using game theory and game experiments [73,74] using or not using diagnostic metrics (such as validation, confusion matrices, ROC analysis, analysis of errors, etc.) [75,76].
The implemented method provides a solution to the calculation of the p-values associated with Anderson–Darling statistics, giving proper weight to the sample size of the investigated experimental data. The advantage of the proposed estimation method, Equation (5), is its very low residual (unexplained variance) and its very high estimation accuracy at convergence (with increasing of in and for p near 1). The main disadvantage is related to its out of range p-values for small AD values, but an extensive simulation study could solve this issue. The worst performances of the implemented methods are observed when simultaneously n is very low (2 or 3) and p is near 0.5 (50–50%).

Author Contributions

L.J. and S.D.B. conceived and designed the experiments; L.J. performed the experiments; L.J. and S.D.B. analyzed the data; S.D.B. wrote the paper and L.J. critically reviewed the manuscript.

Acknowledgments

No grants have been received in support of the research work reported in this manuscript. No funds were received for covering the costs to publish in open access.

Conflicts of Interest

The authors declare no conflict of interest.

References

  1. Nimon, K.F. Statistical assumptions of substantive analyses across the General Linear model: A Mini-Review. Front. Psychol. 2012, 3, 322. [Google Scholar] [CrossRef] [PubMed]
  2. Hoekstra, R.; Kiers, H.A.; Johnson, A. Are assumptions of well-known statistical techniques checked, and why (not)? Front. Psychol. 2012, 3, 137. [Google Scholar] [CrossRef] [PubMed]
  3. Kolmogorov, A. Sulla determinazione empirica di una legge di distribuzione. Giornale dell’Istituto Italiano degli Attuari 1933, 4, 83–91. [Google Scholar]
  4. Smirnov, N. Table for estimating the goodness of fit of empirical distributions. Ann. Math. Stat. 1948, 19, 279–281. [Google Scholar] [CrossRef]
  5. Anderson, T.W.; Darling, D.A. Asymptotic theory of certain “goodness-of-fit” criteria based on stochastic processes. Ann. Math. Stat. 1952, 23, 193–212. [Google Scholar] [CrossRef]
  6. Anderson, T.W.; Darling, D.A. A Test of Goodness-of-Fit. J. Am. Stat. Assoc. 1954, 49, 765–769. [Google Scholar] [CrossRef]
  7. Pearson, K. Contribution to the mathematical theory of evolution. II. Skew variation in homogenous material. Philos. Trans. R. Soc. Lond. 1895, 91, 343–414. [Google Scholar] [CrossRef]
  8. Cramér, H. On the composition of elementary errors. Scand. Actuar. J. 1928, 1, 13–74. [Google Scholar] [CrossRef]
  9. Von Mises, R.E. Wahrscheinlichkeit, Statistik und Wahrheit; Julius Springer: Berlin, Germany, 1928. [Google Scholar]
  10. Shapiro, S.S.; Wilk, M.B. An analysis of variance test for normality (complete samples). Biometrika 1965, 52, 591–611. [Google Scholar] [CrossRef]
  11. Jarque, C.M.; Bera, A.K. Efficient tests for normality, homoscedasticity and serial independence of regression residuals. Econ. Lett. 1980, 6, 255–259. [Google Scholar] [CrossRef]
  12. Jarque, C.M.; Bera, A.K. Efficient tests for normality, homoscedasticity and serial independence of regression residuals: Monte Carlo evidence. Econ. Lett. 1981, 7, 313–318. [Google Scholar] [CrossRef]
  13. Jarque, C.M.; Bera, A.K. A test for normality of observations and regression residuals. Int. Stat. Rev. 1987, 55, 163–172. [Google Scholar] [CrossRef]
  14. D’Agostino, R.B.; Belanger, A.; D’Agostino, R.B., Jr. A suggestion for using powerful and informative tests of normality. Am. Stat. 1990, 44, 316–321. [Google Scholar] [CrossRef]
  15. Lilliefors, H.W. On the Kolmogorov-Smirnov test for normality with mean and variance unknown. J. Am. Stat. Assoc. 1967, 62, 399–402. [Google Scholar] [CrossRef]
  16. Van Soest, J. Some experimental results concerning tests of normality. Stat. Neerl. 1967, 21, 91–97. [Google Scholar] [CrossRef]
  17. Jänstchi, L.; Bolboacă, S.D. Performances of Shannon’s entropy statistic in assessment of distribution of data. Ovidius Univ. Ann. Chem. 2017, 28, 30–42. [Google Scholar] [CrossRef]
  18. Noughabi, H.A. Two Powerful Tests for Normality. Ann. Data Sci. 2016, 3, 225–234. [Google Scholar] [CrossRef]
  19. Zamanzade, E.; Arghami, N.R. Testing normality based on new entropy estimators. J. Stat. Comput. Simul. 2012, 82, 1701–1713. [Google Scholar] [CrossRef]
  20. Peng, H.; Tan, F. Jackknife empirical likelihood goodness-of-fit tests for U-statistics based general estimating equations. Bernoulli 2018, 24, 449–464. [Google Scholar] [CrossRef]
  21. Shah, R.D.; Bühlmann, P. Goodness-of-fit tests for high dimensional linear models. Journal of the Royal Statistical Society. Ser. B Stat. Methodol. 2018, 80, 113–135. [Google Scholar] [CrossRef]
  22. Balakrishnan, K.; Sooriyarachchi, M.R. A goodness of fit test for multilevel survival data. Commun. Stat. Simul. Comput. 2018, 47, 30–47. [Google Scholar] [CrossRef]
  23. Perera, A.A.P.N.M.; Sooriyarachchi, M.R.; Wickramasuriya, S.L. A Goodness of Fit Test for the Multilevel Logistic Model. Commun. Stat. Simul. Comput. 2016, 45, 643–659. [Google Scholar] [CrossRef]
  24. Villaseñor, J.A.; González-Estrada, E.; Ochoa, A. On Testing the inverse Gaussian distribution hypothesis. Sankhya B 2017. [CrossRef]
  25. MacKenzie, D.W. Applying the Anderson-Darling test to suicide clusters: Evidence of contagion at U. S. Universities? Crisis 2013, 34, 434–437. [Google Scholar] [CrossRef] [PubMed]
  26. Müller, C.; Kloft, H. Parameter estimation with the Anderson-Darling test on experiments on glass. Stahlbau 2015, 84, 229–240. [Google Scholar] [CrossRef]
  27. İçen, D.; Bacanlı, S. Hypothesis testing for the mean of inverse Gaussian distribution using α-cuts. Soft Comput. 2015, 19, 113–119. [Google Scholar] [CrossRef]
  28. Ghasemi, A.; Zahediasl, S. Normality tests for statistical analysis: A guide for non-statisticians. Int. J. Endocrinol. Metab. 2012, 10, 486–489. [Google Scholar] [CrossRef] [PubMed]
  29. Hwe, E.K.; Mohd Yusoh, Z.I. Validation guideline for small scale dataset classification result in medical domain. Adv. Intell. Syst. Comput. 2018, 734, 272–281. [Google Scholar] [CrossRef]
  30. Ruxton, G.D.; Wilkinson, D.M.; Neuhäuser, M. Advice on testing the null hypothesis that a sample is drawn from a normal distribution. Anim. Behav. 2015, 107, 249–252. [Google Scholar] [CrossRef]
  31. Lang, T.A.; Altman, D.G. Basic statistical reporting for articles published in biomedical journals: The “Statistical Analyses and Methods in the Published Literature” or The SAMPL Guidelines. In Science Editors’ Handbook; European Association of Science Editors, Smart, P., Maisonneuve, H., Polderman, A., Eds.; EASE: Paris, France, 2013; Available online: http://www.equator-network.org/wp-content/uploads/2013/07/SAMPL-Guidelines-6-27-13.pdf (accessed on 3 January 2018).
  32. Curran-Everett, D.; Benos, D.J. American Physiological Society. Guidelines for reporting statistics in journals published by the American Physiological Society.
  33. Curran-Everett, D.; Benos, D.J. Guidelines for reporting statistics in journals published by the American Physiological Society: The sequel. Adv. Physiol. Educ. 2007, 31, 295–298. [Google Scholar] [CrossRef] [PubMed]
  34. Razali, N.M.; Wah, Y.B. Power comparison of Shapiro-Wilk, Kolmogorov-Smirnov, Lilliefors and Anderson-Darling tests. J. Stat. Model. Anal. 2011, 2, 21–33. [Google Scholar]
  35. Tui, I. Normality Testing—A New Direction. Int. J. Bus. Soc. Sci. 2011, 2, 115–118. [Google Scholar]
  36. Saculinggan, M.; Balase, E.A. Empirical Power Comparison of Goodness of Fit Tests for Normality in the Presence of Outliers. J. Phys. Conf. Ser. 2013, 435, 012041. [Google Scholar] [CrossRef]
  37. Sánchez-Espigares, J.A.; Grima, P.; Marco-Almagro, L. Visualizing type II error in normality tests. Am. Stat. 2017. [CrossRef]
  38. Yap, B.W.; Sim, S.H. Comparisons of various types of normality tests. J. Stat. Comput. Simul. 2011, 81, 2141–2155. [Google Scholar] [CrossRef]
  39. Patrício, M.; Ferreira, F.; Oliveiros, B.; Caramelo, F. Comparing the performance of normality tests with ROC analysis and confidence intervals. Commun. Stat. Simul. Comput. 2017, 46, 7535–7551. [Google Scholar] [CrossRef]
  40. Mbah, A.K.; Paothong, A. Shapiro-Francia test compared to other normality test using expected p-value. J. Stat. Comput. Simul. 2015, 85, 3002–3016. [Google Scholar] [CrossRef]
  41. Arshad, M.; Rasool, M.T.; Ahmad, M.I. Anderson Darling and Modified Anderson Darling Tests for Generalized Pareto Distribution. Pak. J. Appl. Sci. 2003, 3, 85–88. [Google Scholar]
  42. Stephens, M.A. Goodness of fit for the extreme value distribution. Biometrika 1977, 64, 585–588. [Google Scholar] [CrossRef]
  43. D’Agostino, R.B.; Stephens, M.A. Goodness-of-Fit Techniques; Marcel-Dekker: New York, NY, USA, 1986; pp. 123, 146. [Google Scholar]
  44. Shin, H.; Jung, Y.; Jeong, C.; Heo, J.-H. Assessment of modified Anderson–Darling test statistics for the generalized extreme value and generalized logistic distributions. Stoch. Environ. Res. Risk Assess. 2012, 26, 105–114. [Google Scholar] [CrossRef]
  45. De Micheaux, P.L.; Tran, V.A. PoweR: A Reproducible Research Tool to Ease Monte Carlo Power Simulation Studies for Goodness-of-fit Tests in R. J. Stat. Softw. 2016, 69. Available online: https://www.jstatsoft.org/article/view/v069i03 (accessed on 10 April 2018).
  46. 6ixSigma.org—Anderson Darling Test. Available online: http://6ixsigma.org/SharedFiles/Download.aspx?pageid=14&mid=35&fileid=147 (accessed on 2 June 2017).
  47. Spcforexcel. Anderson-Darling Test for Normality. 2011. Available online: http://www.spcforexcel.com/knowledge/basic-statistics/anderson-darling-test-for-normality (accessed on 2 June 2017).
  48. Qimacros—Data Normality Tests Using p and Critical Values in QI Macros. © 2015 KnowWare International Inc. Available online: http://www.qimacros.com/hypothesis-testing//data-normality-test/#anderson (accessed on 2 June 2017).
  49. Jäntschi, L.; Bolboacă, S.D. Distribution Fitting 2. Pearson-Fisher, Kolmogorov-Smirnov, Anderson-Darling, Wilks-Shapiro, Kramer-von-Misses and Jarque-Bera statistics. Bull. Univ. Agric. Sci. Vet. Med. Cluj-Napoca Hortic. 2009, 66, 691–697. [Google Scholar]
  50. Mosteller, F. Questions and Answers—Combining independent tests of significance. Am. Stat. 1948, 2, 30–31. [Google Scholar] [CrossRef]
  51. Bolboacă, S.D.; Jäntschi, L.; Sestraş, A.F.; Sestraş, R.E.; Pamfil, D.C. Pearson-Fisher Chi-Square Statistic Revisited. Information 2011, 2, 528–545. [Google Scholar] [CrossRef]
  52. Rahman, M.; Pearson, L.M.; Heien, H.C. A Modified Anderson-Darling Test for Uniformity. Bull. Malays. Math. Sci. Soc. 2006, 29, 11–16. [Google Scholar]
  53. Matsumoto, M.; Nishimura, T. Mersenne twister: A 623-dimensionally equidistributed uniform pseudo-random number generator (PDF). ACM Trans. Model. Comput. Simul. 1998, 8, 3–30. [Google Scholar] [CrossRef]
  54. Ciocan, A.; Ciocan, R.A.; Gherman, C.D.; Bolboacă, S.D. Evaluation of Patients with Lower Extremity Peripheral Artery Disease by Walking Tests: A Pilot Study. Not. Sci. Biol. 2017, 9, 473–479. [Google Scholar] [CrossRef]
  55. Răcătăianu, N.; Bolboacă, S.D.; Sitar-Tăut, A.-V.; Marza, S.; Moga, D.; Valea, A.; Ghervan, C. The effect of Metformin treatment in obese insulin-resistant patients with euthyroid goiter. Acta Clin. Belg. Int. J. Clin. Lab. Med. 2018. [CrossRef] [PubMed]
  56. Hășmășanu, M.G.; Baizat, M.; Procopciuc, L.M.; Blaga, L.; Văleanu, M.A.; Drugan, T.C.; Zaharie, G.C.; Bolboacă, S.D. Serum levels and ApaI polymorphism of insulin-like growth factor 2 on intrauterine growth restriction infants. J. Matern.-Fetal Neonatal Med. 2018, 31, 1470–1476. [Google Scholar] [CrossRef] [PubMed]
  57. Ciocan, R.A.; Drugan, C.; Gherman, C.D.; Cătană, C.-S.; Ciocan, A.; Drugan, T.C.; Bolboacă, S.D. Evaluation of Chitotriosidase as a Marker of Inflammatory Status in Critical Limb Ischemia. Ann. Clin. Lab. Sci. 2017, 47, 713–719. [Google Scholar] [PubMed]
  58. Bulboacă, A.E.; Bolboacă, S.D.; Stănescu, I.C.; Sfrângeu, C.-A.; Bulboacă, A.C. Preemptive Analgesic and Anti-Oxidative Effect of Curcumin for Experimental Migraine. BioMed Res. Int. 2017, 2017, 4754701. [Google Scholar] [CrossRef]
  59. Bulboacă, A.E.; Bolboacă, S.D.; Bulboacă, A.C.; Prodan, C.I. Association between low thyroid-stimulating hormone, posterior cortical atrophy and nitro-oxidative stress in elderly patients with cognitive dysfunction. Arch. Med. Sci. 2017, 13, 1160–1167. [Google Scholar] [CrossRef] [PubMed]
  60. Nistor, D.-V.; Caterev, S.; Bolboacă, S.D.; Cosma, D.; Lucaciu, D.O.G.; Todor, A. Transitioning to the direct anterior approach in total hip arthroplasty. Is it a true muscle sparing approach when performed by a low volume hip replacement surgeon? Int. Orthopt. 2017, 41, 2245–2252. [Google Scholar] [CrossRef] [PubMed]
  61. Bolboacă, S.D.; Jäntschi, L. Comparison of QSAR Performances on Carboquinone Derivatives. Sci. World J. 2009, 9, 1148–1166. [Google Scholar] [CrossRef] [PubMed]
  62. Harsa, A.M.; Harsa, T.E.; Bolboacă, S.D.; Diudea, M.V. QSAR in Flavonoids by Similarity Cluster Prediction. Curr. Comput.-Aided Drug Des. 2014, 10, 115–128. [Google Scholar] [CrossRef] [PubMed]
  63. Jäntschi, L.; Bolboacă, S.D.; Sestraş, R.E. A Study of Genetic Algorithm Evolution on the Lipophilicity of Polychlorinated Biphenyls. Chem. Biodivers. 2010, 7, 1978–1989. [Google Scholar] [CrossRef] [PubMed]
  64. Chirilă, M.; Bolboacă, S.D. Clinical efficiency of quadrivalent HPV (types 6/11/16/18) vaccine in patients with recurrent respiratory papillomatosis. Eur. Arch. Oto-Rhino-Laryngol. 2014, 271, 1135–1142. [Google Scholar] [CrossRef] [PubMed]
  65. Lenghel, L.M.; Botar-Jid, C.; Bolboacă, S.D.; Ciortea, C.; Vasilescu, D.; Băciuț, G.; Dudea, S.M. Comparative study of three sonoelastographic scores for differentiation between benign and malignant cervical lymph nodes. Eur. J. Radiol. 2015, 84, 1075–1082. [Google Scholar] [CrossRef] [PubMed]
  66. Bolboacă, S.D.; Jäntschi, L. Nano-quantitative structure-property relationship modeling on C42 fullerene isomers. J. Chem. 2016, 2016, 1791756. [Google Scholar] [CrossRef]
  67. Botar-Jid, C.; Cosgarea, R.; Bolboacă, S.D.; Șenilă, S.; Lenghel, M.L.; Rogojan, L.; Dudea, S.M. Assessment of Cutaneous Melanoma by Use of Very- High-Frequency Ultrasound and Real-Time Elastography. Am. J. Roentgenol. 2016, 206, 699–704. [Google Scholar] [CrossRef] [PubMed]
  68. Jäntschi, L.; Balint, D.; Pruteanu, L.L.; Bolboacă, S.D. Elemental factorial study on one-cage pentagonal face nanostructure congeners. Mater. Discov. 2016, 5, 14–21. [Google Scholar] [CrossRef]
  69. Micu, M.C.; Micu, R.; Surd, S.; Girlovanu, M.; Bolboacă, S.D.; Ostensen, M. TNF-a inhibitors do not impair sperm quality in males with ankylosing spondylitis after short-term or long-term treatment. Rheumatology 2014, 53, 1250–1255. [Google Scholar] [CrossRef] [PubMed]
  70. Sestraş, R.E.; Jäntschi, L.; Bolboacă, S.D. Poisson Parameters of Antimicrobial Activity: A Quantitative Structure-Activity Approach. Int. J. Mol. Sci. 2012, 13, 5207–5229. [Google Scholar] [CrossRef] [PubMed]
  71. Bolboacă, S.D.; Jäntschi, L.; Baciu, A.D.; Sestraş, R.E. Griffing’s Experimental Method II: Step-By-Step Descriptive and Inferential Analysis of Variances. JP J. Biostat. 2011, 6, 31–52. [Google Scholar]
  72. EasyFit. MathWave Technologies. Available online: http://www.mathwave.com (accessed on 25 March 2018).
  73. Arena, P.; Fazzino, S.; Fortuna, L.; Maniscalco, P. Game theory and non-linear dynamics: The Parrondo Paradox case study. Chaos Solitons Fractals 2003, 17, 545–555. [Google Scholar] [CrossRef]
  74. Ergün, S.; Aydoğan, T.; Alparslan Gök, S.Z. A Study on Performance Evaluation of Some Routing Algorithms Modeled by Game Theory Approach. AKU J. Sci. Eng. 2016, 16, 170–176. [Google Scholar]
  75. Hossin, M.; Sulaiman, M.N. A review on evaluation metrics for data classification evaluations. Int. J. Data Min. Knowl. Manag. Process 2015, 5, 1–11. [Google Scholar] [CrossRef]
  76. Gopalakrishna, A.K.; Ozcelebi, T.; Liotta, A.; Lukkien, J.J. Relevance as a Metric for Evaluating Machine Learning Algorithms. In Machine Learning and Data Mining in Pattern Recognition; Perner, P., Ed.; Lecture Notes in Computer Science; Springer: Berlin/Heidelberg, Germany, 2013; Volume 7988. [Google Scholar]
Figure 1. Probability as function of the AD statistic for a selected case (n = 25) in the Monte Carlo experiment: (a) p = p(AD); (b) p = p(eAD); (c) α-1 vs. eAD; (d) −ln(α) vs. AD.
Figure 1. Probability as function of the AD statistic for a selected case (n = 25) in the Monte Carlo experiment: (a) p = p(AD); (b) p = p(eAD); (c) α-1 vs. eAD; (d) −ln(α) vs. AD.
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Figure 2. The effect in differences between classical and stratified random in calculated AD statistic.
Figure 2. The effect in differences between classical and stratified random in calculated AD statistic.
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Figure 3. Distribution of residuals (differences between MC-simulated values and the values estimated by our model) for the probability from regression for the whole pool of data (30,000 pairs). (a) untransformed data (b) log transformed data
Figure 3. Distribution of residuals (differences between MC-simulated values and the values estimated by our model) for the probability from regression for the whole pool of data (30,000 pairs). (a) untransformed data (b) log transformed data
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Figure 4. 3D plot of the estimation error for data expressed in logarithm scale as function of p (ranging from 0.500 to 0.999) and n (ranging from 2 to 61).
Figure 4. 3D plot of the estimation error for data expressed in logarithm scale as function of p (ranging from 0.500 to 0.999) and n (ranging from 2 to 61).
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Figure 5. 3D plot of the estimation error for untransformed data: Z-axis show the 105·(p p ^ ) as a function of p (ranging from 0.500 to 0.999) and n (ranging from 2 to 61).
Figure 5. 3D plot of the estimation error for untransformed data: Z-axis show the 105·(p p ^ ) as a function of p (ranging from 0.500 to 0.999) and n (ranging from 2 to 61).
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Figure 6. Normal probability plots (P–P) and quantile-quantile plot (Q–Q) by example: graphs for set 9 (n = 70) in the first row, and for set 11 (n = 40) in the second row.
Figure 6. Normal probability plots (P–P) and quantile-quantile plot (Q–Q) by example: graphs for set 9 (n = 70) in the first row, and for set 11 (n = 40) in the second row.
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Table 1. The goodness-of-fit tests: approaches.
Table 1. The goodness-of-fit tests: approaches.
Test NameAbbreviationProcedure
Kolmogorov–SmirnovKSProximity analysis of the empirical distribution function (obtained on the sample) and the hypothesized distribution (theoretical)
Anderson–DarlingADHow close the points are to the straight line estimated in a probability graphic
chi-squareCSComparison of sample data distribution with a theoretical distribution
Cramér–von MisesCMEstimation of the minimum distance between theoretical and sample probability distribution
Shapiro–WilkSWBased on a linear model between the ordered observations and the expected values of the ordered statistics of the standard normal distribution
Jarque–BeraJBEstimation of the difference between asymmetry and kurtosis of observed data and theoretical distribution
D’Agostino–PearsonAPCombination of asymmetry and kurtosis measures
LillieforsLFA modified KS that uses a Monte Carlo technique to calculate an approximation of the sampling distribution
Table 2. Anderson–Darling test: critical values according to significance level.
Table 2. Anderson–Darling test: critical values according to significance level.
Distribution [Ref]α = 0.10α = 0.05α = 0.01
Normal & lognormal [43]0.6310.7521.035
Weibull [43]0.6370.7571.038
Generalized extreme value [44]---
n = 100.2360.2760.370
n = 200.2320.2740.375
n = 300.2320.2760.379
n = 400.2330.2770.381
n = 500.2330.2770.383
n = 1000.2340.2790.387
Generalized logistic [44]---
n = 100.2230.2660.374
n = 200.2410.2900.413
n = 300.2200.3010.429
n = 400.2540.3060.435
n = 500.2580.3110.442
n = 1000.2670.3230.461
Uniform [52] *1.9362.4993.903
* Expressed as upper tail percentiles.
Table 3. Anderson–Darling for small sizes: p-values formulas.
Table 3. Anderson–Darling for small sizes: p-values formulas.
Anderson–Darling StatisticFormula for p-Value Calculation
AD ≥ 0.6exp (1.2937 − 5.709∙(AD*) + 0.0186∙(AD*)2)
0.34 < AD* < 0.6exp (0.9177 − 4.279∙(AD*) − 1.38∙(AD*)2)
0.2 < AD* < 0.341 − exp (−8.318 + 42.796∙(AD*) − 59.938∙(AD*)2)
AD* ≤ 0.21 − exp (−13.436 + 101.14∙(AD*) − 223.73∙(AD*)2)
A D * = A D ( 1 + 0.75 n + 2.25 n 2 ) ; A D = n 1 n · i = 0 n ( 2 · i 1 ) · [ ln ( F ( X i ) + ln ( 1 F ( X n i + 1 ) ) ] .
Table 4. Cases for the half-split of [0, 1).
Table 4. Cases for the half-split of [0, 1).
Classt1t2t3Case
“0” if ti < 0.5
“1” if ti ≥ 0.5
0001
0012
0103
0114
1005
1016
1107
1118
Table 5. Unique cases for the half-split of [0, 1).
Table 5. Unique cases for the half-split of [0, 1).
|{ti|ti < 0.5}||{ti|ti ≥ 0.5}|Frequency (Case in Table 4)
301 (case 1)
213 (case 2, 3, 5)
123 (case 4, 6, 7)
031 (case 8)
Table 6. Proposed model tested for the AD = AD(p) series for n = 25. SST: Sum of Squares: Total; SSRes: Sum of Squares: Residuals; SSE = Sum of Squares Error.
Table 6. Proposed model tested for the AD = AD(p) series for n = 25. SST: Sum of Squares: Total; SSRes: Sum of Squares: Residuals; SSE = Sum of Squares Error.
CoefficientValue (95% CI)SEt-Value
a04.160 (4.126 to 4.195)0.017567237
a1−10.327 (−10.392 to −10.263)0.032902−314
a29.357 (9.315 to 9.400)0.02178430
a3−6.147 (−6.159 to −6.135)0.00601−1023
a43.4925 (3.4913 to 3.4936)0.0005835993
SST = 1550651, SSRes = 0.0057,
SSE = 0.0034, r2adj = 0.999999997
Table 7. Coefficients of the proposed model and their Student t-values provided in round brackets.
Table 7. Coefficients of the proposed model and their Student t-values provided in round brackets.
bi,j (ti,j)j = 0j = 1j = 2j = 3j = 4
i = 05.6737
(710)
−38.9087
(4871)
88.7461
(11111)
−179.5470
(22479)
199.3247
(24955)
i = 1−13.5729
(1699)
83.6500
(10473)
−181.6768
(22746)
347.6606
(43526)
−367.4883
(46009)
i = 212.0750
(1512)
−70.3770
(8811)
139.8035
(17503)
−245.6051
(30749)
243.5784
(30496)
i = 3−7.3190
(916)
30.4792
(3816)
−49.9105
(6249)
76.7476
(9609)
−70.1764
(8786)
i = 43.7309
(467)
−6.1885
(775)
7.3420
(919)
−9.3021
(1165)
7.7018
(964)
Table 8. Residuals: descriptive statistics.
Table 8. Residuals: descriptive statistics.
Parameter(p p ^ )ln(p p ^ )log(p p ^ )
Arithmetic mean3.04 × 10−7−18.8283−8.17703
Standard deviation2.55 × 10−63.94771.7144
Standard error1.47 × 10−80.022790.009898
Median1.5 × 10−8−18.0132−7.82304
Mode9.52 × 10−8−16.1677−7.02156
Minimum1.32 × 10−18−41.167−17.8786
Maximum0.000121−9.02296−3.9186
Table 9. Characteristics of the investigated datasets.
Table 9. Characteristics of the investigated datasets.
Set IDWhat the Data Represent?Sample SizeReference
1Distance (m) on treadmill test, applied on subject ts with peripheral arterial disease24[54]
2Waist/hip ratio, determined in obese insulin-resistant patients53[55]
3Insulin-like growth factor 2 (pg/mL) on newborns60[56]
4Chitotriosidase activity (nmol/mL/h) on patients with critical limb ischemia43[57]
5Chitotriosidase activity (nmol/mL/h) on patients with critical limb ischemia and on controls86[57]
6Total antioxidative capacity (Eq/L) on the control group10[58]
7Total antioxidative capacity (Eq/L) on the group with induced migraine40[53]
8Mini mental state examination score (points) elderly patients with cognitive dysfunction163[59]
9Myoglobin difference (ng/mL) (postoperative–preoperative) in patients with total hip arthroplasty70[60]
10The inverse of the molar concentration of carboquinone derivatives, expressed in logarithmic scale37[61]
11Partition coefficient expressed in the logarithmic scale of flavonoids40[62]
12Evolution of determination coefficient in the identification of optimal model for lipophilicity of polychlorinated biphenyls using a genetic algorithm30[63]
13Follow-up days in the assessment of the clinical efficiency of a vaccine 31[64]
14Strain ratio elastography to cervical lymph nodes50[65]
15Total strain energy (eV) of C42 fullerene isomers45[66]
16Breslow index (mm) of melanoma lesions29 [67]
17Determination coefficient distribution in full factorial analysis on one-cage pentagonal face C40 congeners: dipole moment44[68]
18The concentration of spermatozoids (millions/mL) in males with ankylosing spondylitis60[69]
19The parameter of the Poisson distribution31[70]
20Corolla diameter of Calendula officinalis L. for Bon-Bon Mix × Bon-Bon Orange28[71]
Table 10. Anderson–Darling (AD) statistic, associated p-values, and test conclusion: comparisons.
Table 10. Anderson–Darling (AD) statistic, associated p-values, and test conclusion: comparisons.
SetEasyFitOur MethodSPC for Excel
AD StatisticReject H0?p-ValueReject H0?p-ValueReject H0?
11.18No0.2730No0.0035Yes
21.34No0.2198No0.0016Yes
315.83Yes3.81 × 10−8Yes0.0000Yes
41.59No0.1566No4.63 × 10−15Yes
56.71Yes0.0005Yes1.44 × 10−16Yes
60.18Noo.o.r. 0.8857No
73.71Yes0.0122Yes1.93 × 10−9Yes
811.70Yes2.49 × 10−6Yes3.45 × 10−28Yes
90.82No0.4658No0.0322Yes
100.60No0.6583No0.1109No
110.81No0.4752No0.0334Yes
120.34Noo.o.r. 0.4814No
134.64Yes0.0044Yes0.0000Yes
141.90No0.1051No0.0001Yes
150.39No0.9297No0.3732No
160.67No0.5863No0.0666No
175.33Yes0.0020Yes2.23 × 10−13Yes
182.25No0.0677No9.18 × 10−6Yes
191.30No0.2333No0.0019Yes
200.58No0.6774No0.1170No
AD = Anderson–Darling; o.o.r = out of range.
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