Mathematical Problems in Engineering

Volume 2008, Article ID 186372, 12 pages

http://dx.doi.org/10.1155/2008/186372

## A Mathematical Tool for Inference in Logistic Regression with Small-Sized Data Sets: A Practical Application on ISW-Ridge Relationships

^{1}Department of Business Administration, Shu-Te University, Yen Chau, Kaohsiung, Taiwan 82445, Taiwan^{2}Department of Management Information System, Yung-Ta Institute of Technology and Commerce, Pingtung, Taiwan 90941, Taiwan^{3}Department of Risk Management and Insurance, Kaohsiung First University of Science and Technology, Kaohsiung, Taiwan 811, Taiwan^{4}Department of Logistics Management, Shu-Te University, Yen Chau, Kaohsiung, Taiwan 82445, Taiwan

Received 6 October 2007; Revised 3 March 2008; Accepted 26 August 2008

Academic Editor: Irina Trendafilova

Copyright © 2008 Tsung-Hao Chen et al. This is an open access article distributed under the Creative Commons Attribution License, which permits unrestricted use, distribution, and reproduction in any medium, provided the original work is properly cited.

#### Abstract

The general approach to modeling binary data for the purpose of estimating the propagation of an internal solitary wave (ISW) is based on the maximum likelihood estimate (MLE) method. In cases where the number of observations in the data is small, any inferences made based on the asymptotic distribution of changes in the deviance may be unreliable for binary data (the model's lack of fit is described in terms of a quantity known as the deviance). The deviance for the binary data is given by D. Collett (2003). may be unreliable for binary data. Logistic regression shows that the -values for the likelihood ratio test and the score test are both 0.05. However, the null hypothesis is not rejected in the Wald test. The seeming discrepancies in -values obtained between the Wald test and the other two tests are a sign that the large-sample approximation is not stable. We find that the parameters and the odds ratio estimates obtained via conditional exact logistic regression are different from those obtained via unconditional asymptotic logistic regression. Using exact results is a good idea when the sample size is small and the approximate -values are 0.10. Thus in this study exact analysis is more appropriate.

#### 1. Introduction

Internal waves refer to the motion at the interface between layers of water of different densities in a stratified water body, such as the ocean. The simplest oceanic density structure, where differences in water density are mostly caused by differences in water temperature or salinity, can be approximated by a two-layer model. Oceanic internal waves typically have wavelengths ranging from hundreds of meters to tens of kilometers, with periods from tens of minutes to tens of hours. In the Andaman and Sulu Sea they can have amplitudes (peak to trough distance) exceeding 50m and in the South China Sea the amplitude can exceed 110 m [1–9]. The mixing and dissipation generated by internal waves have important effects on the cross slope exchange processes, enhancement of bottom stress, and generation of the nepheloid layers. It has recently been proposed that internal waves may make a significant contribution to internal oceanic mixing and hence have an important influence on climatic change. This is why it is necessary to scrutinize the interaction of nonlinear internal solitary waves (ISWs) with the seabed topography [10–17].

Several studies, including both simulations and laboratory experiments, aiming at exploring the mechanisms for the generation, propagation, and evolution of ISWs, have already been carried out. However, since energy dissipation plays such an important and varied role on water and sedimentary movement in coastal seas [18], we need a better fitting and more appropriate model for predicting ISW propagation. A preliminary approach has recently been made in which the effects of weighted parameters on the amplitude and reflection of energy-based ISWs from uniform slopes in a two layered fluid system were investigated [19]. The results are quite consistent with other experimental results, and are applicable to the naturally occurring reflection of ISWs from sloping bottoms. More recently, Chen et al. [20] concluded the goodness-of-fit and predictive ability of the cumulative logistic regression models to be better than that of the binary logistic regression models. However, in cases where the data are so small that there are some observations with proportions close to zero or one, inferences based on the asymptotic distribution of the change in deviance may be unreliable. In point of fact, reports on statistical manipulations related to this theme are rather rare.

The rest
of the paper is organized as follows. In Section 2 we describe the experimental
set-up and theoretical background needed to understand the hydrodynamic
interaction. We also discuss the analysis of the logistic regression
model, and introduce the exact conditional logistic model and the hypothesis on which the parameters are based.
Section
3 is devoted to a comparison of the conditional exact logistic regression model and the
unconditional asymptotic logistic regression model. Finally, some conclusions
are made. It is noted that small sample size
means that there are some observations with proportions close to zero or one
and *P*-values of less than 0.10, which
is an indication that an exact analysis would be more appropriate.

#### 2. Research Framework

Experiments were carried out in the laboratory using a
two-layer fluid system of fresh and briny water in a 12 m long wave flume (rectangular in
cross-section). The
upper layer of water in the wave
flume consisted of fresh water with a density and a depth , while the lower layer was comprised of
brine with a density and
a depth . The leading ISW was generated by the lifting of a
pneumatic sluice gate at one end of the flume. The wave propagated into the
main section of the flume to the left-hand side (LHS) of the gate. The amplitude *a* and characteristic length of the ISW were predetermined by arranging the step
length *L* and step depth (see Figure 1). Six ultrasonic probes connected to an amplifier
unit and A/D converter, then to a personal computer, gathered and processed
digital signals as the ISW propagated along the flume. As
the ISW propagated from the RHS (right-hand side) to the LHS of the
flume, the first ultrasonic probe (P1) recorded the properties of the incident
ISW, the wave amplitude and characteristic length, while the second probe (P2)
collected reflected signals showing the wave-obstacle interaction. The methodology for
measuring the physical properties related to the propagation and dissipation of
the ISW has been reported in detail by Chen
et al. [21]. The
amplitude-based transmission rate during the wave-ridge interaction was
dependent on two factors, ridge height and potential energy.

##### 2.1. Exact Conditional Logistic Regression Model

The theoretical basis for the exact conditional logistic regression model was originally laid down by Cox [22], but recent algorithmic advances in computing the exact distributions have made the methodology more practical. Since then Hirji et al. [23] have developed an efficient algorithm for generating the required conditional distributions. Cox and Snell [24] noted that it has been known since the 1970’s how to extend the theory of Fisher’s exact test to logistic regression models. The interested reader may refer to Mehta and Patel [25] for a useful summary of exact logistic regression. A complete discussion of the exact logistic regression methodology and more detailed applications can be found in a variety of sources [26–30].

Here, let represent the probability of “success” for a binary response for the explanatory variables . The notation can be simplified by using to represent the conditional mean of given when a logistic distribution is utilized: such that The transformation of , which is central to this study of logistic regression, is the logit transformation. This transformation is defined as where is an unknown parameter vector.

The sufficient statistics for the in the unconditional likelihood function are where is the realization of .

If and indicate sufficient statistics corresponding to and , the conditional probability density function of conditional on can be formulated as where indicates the number of vectors , such that and .

Conditional exact inference involves the generation of the conditional permutational distribution for the sufficient statistics for the parameters. The distribution is called the permutation conditional distribution or exact conditional distribution.

##### 2.2. Testing the Hypotheses

According to exact logistic regression (for both the exact score conditional test and the probability test) the parameters for the specified hypothesis are equal to zero. If an effect consists of two or more parameters, then it is hypothesized that all the parameters are simultaneously equal to zero [26, 27].

###### 2.2.1. Exact Score Conditional Test

The null
hypothesis is The conditional
mean and variance matrix of (conditional on ) are calculated via the exact
conditional scores test. The score statistic is Now
compare this to the score for each member of the distribution In the
null hypothesis, an exact *P*-value,
which is the probability of obtaining a more extreme statistic than the
observed one, is assumed.

The
result of the *P*-value is where A mid *P*-value,
adjusted for the discreteness of the distribution, is assumed for the null
hypothesis.

The mid-*p* statistic is defined as

###### 2.2.2. Probability Testing

For small samples, the parameter inference process is
carried out using conditional distribution probabilities, such as exact *P*-values, rather than a crude approximation
[29]. For testing the null hypothesis we use Under the null hypothesis, the
exact probability test statistic is just ; the corresponding *P*-value
gives the probability of getting a less likely statistic where

#### 3. Analytical Results

The effects of the ridge height, the depth of the lower water layer, and the potential energy on the propagation of the ISW are all considered. The results from the laboratory experiments are shown in the data sets. The amplitudes of the incident and reflected waves are also included. The dependent variables for the binary logistic regression model are classified into two groups, weak and strong, based on the amplitude incident rate. When the hypothetical incident rate is >0.5 it is considered strong and when it is <0.5 it is considered weak. The frequencies for the strong and weak levels are 35 and 28, respectively.

##### 3.1. Asymptotic Logistic Regression Model

The methodologies utilized in the asymptotic logistic regression model and the diagnostics of the goodness-of-fit statistics are discussed below.

###### 3.1.1. Goodness-of-Fit Statistics

The Pearson Chi-squares test and deviance Chi-squares test are used. The results of the Pearson Chi-square test give a distribution with the degrees of freedom , where is the number of explanatory variables, is the number of response levels, and is the number of subpopulations.

The goodness-of-fit
statistics are shown in Table 1. The dispersion parameter (value/DF), which
indicates estimated deviance, is given in the value/DF column. The dispersion
parameter is 0.7268 and the Pearson Chi-squares dispersion parameter is 1.2157.
Ideally, this value should be very close to 1.00. The values of the Pearson Chi-square
and deviance Chi-square statistics are 60.7842 and 36.338, respectively, with
50 degrees of freedom . The Pearson Chi-squares value is
slightly larger than the degrees of freedom; the *P*-values for the deviance and Pearson Chi-squares
are all larger than 0.05 (0.9260, 0.1412). From this we see that although there
is a little over dispersion, this model seems to have an acceptable fit
with the data. The overdispersion means that the model still needs to be modified.

###### 3.1.2. Regression Diagnostics

There are a number of different ways to plot the regression diagnostics, each directed at a particular aspect of the fit. For examples see Hosmer and Lemeshow [28], and Landwehr et al. [31] who discussed graphical techniques for logistic regression diagnostics. Generally such techniques offer a visual rather than numerical representation that may be more intuitively appealing to some researchers. Index plots are useful for the identification of extreme values [32]. An examination of the index plots of the Pearson residuals (Figure 2) and the deviance residuals (Figure 3) for our data indicates that case 11 and case 27 are poorly accounted for by the model. It can be seen in the index plot of the diagonal elements of the hat matrix (Figure 4) that case 49 is at the extreme point in the design space.

###### 3.1.3. Outliers and Influential Observations

The values of outliers can be quite substantial and influential. A look at Table 5 shows the advantage of removing such observations from the data (here, case 11, case 27, and case 49), then refitting the newly revised model to the remaining observations.

The goodness-of-fit statistics are presented in Table 2.
The estimates of deviance are shown in the column marked value/DF. The
dispersion parameter (value/DF) is 0.3376 and the Pearson Chi-square dispersion parameter is 0.4752. The values of the deviance
and Pearson Chi-square are less than the
degrees of freedom, while the *P*-values
of the deviance and Pearson Chi-square are all >0.05 (i.e., 1.0000, 0.9993,
resp.). These indicate that this model seems to have an acceptable fit with the
data.

###### 3.1.4. Testing the Global Null Hypothesis:

When testing the null hypothesis for large samples, the
explanatory variables have coefficients of zero. According to the Chi-squares
analysis, the associated *P*-values are
all approximately zero, suggesting that the explanatory coefficients are all zero.

The results obtained after rerunning
the unconditional asymptotic logistic
regression after the removal
of some of the observations from the data (i.e., case 11, case 27, and case 49) (see
Table 3) still contain some unconditional asymptotic
results. These results are obtained by deriving the Chi-square statistics while
testing for the global null hypothesis
(likelihood ratio, score, and Wald tests). For the likelihood ratio and score
tests, the null hypothesis that is zero is rejected, but not
for the Wald test. The seeming discrepancies in *P*-values obtained between the Wald test and the other two tests are
a sign that the large-sample approximation is not stable.

##### 3.2. Exact Logistic Regression Model

Exact
logistic regression for binary outcomes can be utilized to provide an exact
score test and an exact probability test for hypotheses where the parameters
are equal to zero; these tests produce an exact *P*-value and a mid *P*-value.

To
test whether individual parameter estimates are zero, we also require point
estimates of the parameters, an odds ratio that contains two-sided confidence
limits, and the *P*-value.

###### 3.2.1. Conditional Exact Tests:

The results of exact conditional analysis obtained using the exact logistic regression model are shown in Table 4. The results for the exact score conditional test and the probability test are also reported in this table. For the joint test it is required that all the parameters for the exact statement be simultaneously equal to zero, that is, the null hypothesis is .

In the joint test results an exact *P*-value of <.0001 is produced; the probability
test produces an exact *P-*value of
0.0023. These test results lead to
a rejection that the null hypothesis of is zero. This
shows that the ridge height ,
lower layer water depth , and potential
energy
are significant for the joint exact test.

Given the effects of the ridge height , lower
layer water depth , and potential
energy , the
exact *P-*value and mid *P-*value are both <.0001. These results
lead to a rejection of the
null hypothesis that is
zero.To put
it another way, ridge height ,
lower layer water depth , and potential
energy are significant factors associated with the amplitude-based
incident rate.

###### 3.2.2. Parameter Estimation and Odds Ratio Estimation

Stokes et al. [27] have suggested that large sample
theory may not be appropriate for small-sized
data. This thus means that tests based on the asymptotic normality of the MLEs
may be unreliable. They recommend that when sample sizes are small, with
approximate *P*-values of less than
0.10, it is a good idea to look at the exact results. If the approximate *P*-values are larger than 0.15, then the
approximate methods are probably satisfactory, in the sense that the exact
results are likely to agree with them.

*Parameter Estimates
for Unconditional Asymptotic Logistic Regression*

The analytical results for the estimated maximum
likelihood and odds ratios are shown in Tables
5 and
6. The ridge height , lower layer water depth , and potential energy are all significant
factors affecting the amplitude-based incident rate (*P* = .0106, *P* = .0053, and *P* = .0067, resp.).

The fitted unconditional asymptotic logistic regression
lines can be stated as

*Parameter Estimation for Conditional Exact Logistic
Regression*

The analytical
results of the exact parameter estimates and exact odds ratio estimates are
presented in Tables 7 and 8, respectively. The ridge height , lower layer water depth , and potential energy are all significant
factors affecting the amplitude-based incident rate (*P* < .0001). We create
a median unbiased estimate instead of the conditional MLE, because the value of
the observed sufficient statistic lies at the extreme end of the derived
distribution. The implication is that the conditional MLE does not exist. Even
though the asymptotic results are unreliable, the exact analysis allows us to
conclude that these factors have a significant effect. The fitted conditional
exact logistic regression lines can be formulated as We can see from
Tables 5 and
7 that the parameters obtained from conditional exact logistic
regression are smaller than those obtained from unconditional asymptotic
logistic regression, but the *P*-values
of the unconditional asymptotic estimates are larger than those of the exact
estimates. A comparison of the odds ratio estimates (in
Tables 6 and 8) shows that the parameters obtained from
the conditional exact logistic regression are different than those obtained
from the unconditional asymptotic logistic regression.

Stokes et al. [27] recommended that when sample sizes are small and the
approximate *P*-values are less than
0.10, it is better to look at the exact results. Thus in this study, the small
sample size and *P*-values make exact
analysis more appropriate.

#### 4. Conclusions

A laboratory experiment is designed to investigate the propagation of an internal solitary wave over a submerged ridge. Analytical methods and a logistic regression model are employed to examine the amplitude-based incident rate. Large sample theory may not be suitable for data with small cell counts. This tends to make tests based on the asymptotic normality of the MLEs unreliable.

The ridge height, lower layer water depth, and potential energy are considered in the regression model. Once a model has been fitted to the observed values of a binary response variable, it is essential to check the validity of the fit. We discuss some methods for exploring the adequacy of the model and some diagnostic methods. The techniques used to examine the adequacy of a fitted unconditional asymptotic logistic regression model and conditional exact logistic regressions are known as diagnostics methods for testing the global null hypothesis. Based on the analytical results we can draw the following conclusions.

(1) The unconditional asymptotic logistic model results lead us to the conclusion that the three explanatory variables (ridge height, lower layer water depth, and potential energy) are significant factors affecting the amplitude-based incident rate. Both deviance and Pearson Chi-square tests are used to examine the goodness-of-fit of the model. The dispersion parameter for the estimate of deviance (value/DF) is 0.7268, and the Pearson Chi-square dispersion parameter is 1.2157. Preferably, this value should be very close to 1.00. The Pearson parameter is slightly larger than the degrees of freedom. We note that there is still a little overdispersion with this model which means that it needs to be modified.

(2)
A look at the index plots
for the Pearson residuals (Figure 2) and the deviance residuals (Figure 3)
shows that case 11 and case 27 are poorly accounted for by the model. In the
index plot of the diagonal elements of the hat matrix (Figure 4), case 49 is an
extreme point in the design space. After these observations (case 11, case 27,
and case 49) are removed from the data, the new revised model is refitted based
on the remaining observations. The values of
the deviance and Pearson Chi-squares are now less than the degrees of freedom,
and the *P*-values for deviance and
Pearson Chi-square are all >0.05 (1.0000, 0.9993, resp.). In other words,
this revised model seems to fit the data acceptably well.

(3)
When testing the global null hypothesis , only three
Chi-square statistics (likelihood ratio, score, and Wald tests) are generated. The *P*-values obtained by logistic
regression for the likelihood ratio test and score test are both <0.05.
However, the null hypothesis is not rejected for the Wald test. The seeming
discrepancies in *P*-values obtained
between the Wald test and the other two tests are a sign that the large-sample
approximation is not stable.

(4) The results of exact conditional analysis from the exact logistic regression model are shown in Table 4. The ridge height , lower layer water depth , and potential energy are all significant in the joint results. The ridge height , lower layer water depth , and potential energy effects are all significant factors affecting the amplitude-based incident rate.

(5)
A comparison of the parameters shown in
Tables 6 and 8 and the
odds ratio estimates in Tables 6 and
8
shows that the parameters and the odds ratio estimates obtained from conditional exact logistic regression are
different from those obtained from unconditional asymptotic logistic regression. As recommended by Stokes
et al. [27], in cases of small
sample sizes where the approximate *P*-values
are less than 0.10, it is a good idea to look at the exact results. For this
study, the small sample size and *P*-values
indicate that an exact analysis would be more appropriate.

#### Acknowledgments

The authors would like to thank the National Science Council of the Republic of China, Taiwan for financial support of this research under Contracts no. NSC 96-2628-E-366-004-MY2 and NSC 96-2628-E-132-001-MY2. They also wish to thank the editor of Mathematical Problems in Engineering, and the three anonymous reviewers for their helpful suggestions on the improvement of this paper.

#### References

- A. R. Osborne and T. L. Burch, “Internal solitons in the Andaman Sea,”
*Science*, vol. 208, no. 4443, pp. 451–460, 1980. View at Publisher · View at Google Scholar - J. R. Apel, J. R. Holbrook, J. Tsai, and A. K. Liu, “The Sulu Sea internal soliton experiment,”
*Journal of Physical Oceanography*, vol. 15, no. 12, pp. 1625–1651, 1985. View at Publisher · View at Google Scholar - A. K. Liu, J. R. Holbrook, and J. R. Apel, “Nonlinear internal wave evolution in the Sulu Sea,”
*Journal of Physical Oceanography*, vol. 15, no. 12, pp. 1613–1624, 1985. View at Publisher · View at Google Scholar - A. K. Liu, Y. S. Chang, M. K. Hsu, and N. K. Liang, “Evolution of nonlinear internal waves in the East and South China Seas,”
*Journal of Geophysical Research*, vol. 103, no. C4, pp. 7995–8008, 1998. View at Publisher · View at Google Scholar - A. K. Liu and M. K. Hsu, “Internal wave study in the South China Sea using Synthetic Aperture Radar (SAR),”
*International Journal of Remote Sensing*, vol. 25, no. 7-8, pp. 1261–1264, 2004. View at Publisher · View at Google Scholar - M. K. Hsu and A. K. Liu, “Nonlinear internal waves in the South China Sea,”
*Canadian Journal of Remote Sensing*, vol. 26, no. 2, pp. 72–81, 2000. View at Google Scholar - M. K. Hsu, A. K. Liu, and C. H. Lee, “Using SAR image to study internal waves in the Sulu Sea,”
*Journal of Photogrammetry and Remote Sensing*, vol. 3, pp. 1–14, 2003. View at Google Scholar - K. Zeng and W. Alpers, “Generation of internal solitary waves in the Sulu Sea and their refraction by bottom topography studied by ERS SAR imagery and a numerical model,”
*International Journal of Remote Sensing*, vol. 25, no. 7-8, pp. 1277–1281, 2004. View at Publisher · View at Google Scholar - Q. Zheng, R. D. Susanto, C.-R. Ho, Y. T. Song, and Q. Xu, “Statistical and dynamical analyses of generation mechanisms of solitary internal waves in the northern South China Sea,”
*Journal of Geophysical Research*, vol. 112, no. 3, Article ID C03021, 2007. View at Publisher · View at Google Scholar - T. W. Kao, F.-S. Pan, and D. Renouard, “Internal solitons on the pycnocline: generation, propagation, and shoaling and breaking over a slope,”
*Journal of Fluid Mechanics*, vol. 159, pp. 19–53, 1985. View at Publisher · View at Google Scholar - K. R. Helfrich, “Internal solitary wave breaking and run-up on a uniform slope,”
*Journal of Fluid Mechanics*, vol. 243, pp. 133–154, 1992. View at Publisher · View at Google Scholar - F. Wessels and K. Hutter, “Interaction of internal waves with a topographic sill in a two-layered fluid,”
*Journal of Physical Oceanography*, vol. 26, no. 1, pp. 5–20, 1996. View at Publisher · View at Google Scholar - H. Michallet and G. N. Ivey, “Experiments on mixing due to internal solitary waves breaking on uniform slopes,”
*Journal of Geophysical Research*, vol. 104, no. C6, pp. 13467–13477, 1999. View at Publisher · View at Google Scholar - C.-Y. Chen and J. R.-C. Hsu, “Interaction between internal waves and a permeable seabed,”
*Ocean Engineering*, vol. 32, no. 5-6, pp. 587–621, 2005. View at Publisher · View at Google Scholar - C.-Y. Chen, J. R.-C. Hsu, C.-F. Kuo, H.-H. Chen, and M.-H. Cheng, “Laboratory observations on internal solitary wave evolution over a submarine ridge,”
*China Ocean Engineering*, vol. 20, no. 1, pp. 61–72, 2006. View at Google Scholar - C.-Y. Chen, J. R.-C. Hsu, M.-H. Cheng, H.-H. Chen, and C.-F. Kuo, “An investigation on internal solitary waves in a two-layer fluid: propagation and reflection from steep slopes,”
*Ocean Engineering*, vol. 34, no. 1, pp. 171–184, 2007. View at Publisher · View at Google Scholar - C.-Y. Chen, “An experimental study of stratified mixing caused by internal solitary waves in a two-layered fluid system over variable seabed topography,”
*Ocean Engineering*, vol. 34, no. 14-15, pp. 1995–2008, 2007. View at Publisher · View at Google Scholar - D. A. Cacchione, L. F. Pratson, and A. S. Ogston, “The shaping of continental slopes by internal tides,”
*Science*, vol. 296, no. 5568, pp. 724–727, 2002. View at Publisher · View at Google Scholar - C.-W. Chen, C.-Y. Chen, P. H.-C. Yang, and T.-H. Chen, “Analysis of experimental data on internal waves with statistical method,”
*Engineering Computations*, vol. 24, no. 2, pp. 116–150, 2007. View at Publisher · View at Google Scholar - C.-W. Chen, H.-C. P. Yang, C.-Y. Chen, A. K.-H. Chang, and T.-H. Chen, “Evaluation of inference adequacy in cumulative logistic regression models: an empirical validation of ISW-ridge relationships,”
*China Ocean Engineering*, vol. 22, no. 1, pp. 43–56, 2008. View at Google Scholar - C.-Y. Chen, J. R.-C. Hsu, C.-W. Chen, C.-F. Kuo, H.-H. Chen, and M.-H. Cheng, “Wave propagation at the interface of a two-layer fluid system in the laboratory,”
*Journal of Marine Science and Technology*, vol. 15, no. 1, pp. 8–16, 2007. View at Google Scholar - D. R. Cox,
*The Analysis of Binary Data*, Methuen, London, UK, 1970. View at Zentralblatt MATH · View at MathSciNet - K. F. Hirji, C. R. Mehta, and N. R. Patel, “Computing distributions for exact logistic regression,”
*Journal of the American Statistical Association*, vol. 82, no. 400, pp. 1110–1117, 1987. View at Publisher · View at Google Scholar · View at Zentralblatt MATH · View at MathSciNet - D. R. Cox and E. J. Snell,
*Analysis of Binary Data*, vol. 32 of*Monographs on Statistics and Applied Probability*, Chapman & Hall, London, UK, 2nd edition, 1989. View at Zentralblatt MATH · View at MathSciNet - C. R. Mehta and N. R. Patel, “Exact logistic regression: theory and examples,”
*Statistics in Medicine*, vol. 14, no. 19, pp. 2143–2160, 1995. View at Publisher · View at Google Scholar - R. E. Derr, “Performing exact regression with the SAS system,” in
*Proceedings of the 25th Annual SAS Users Group International Conference*, Cary, NC, USA, April 2000, paper P254-25. - M. E. Stokes, C. S. Davis, and G. G. Koch,
*Categorical Data Analysis Using the SAS System*, SAS Institute, Cary, NC, USA, 2000. - D. W. Hosmer and S. Lemeshow,
*Applied Logistic Regression*, John Wiley & Sons, New York, NY, USA, 2nd edition, 2000. View at Zentralblatt MATH - A. Agresti,
*Categorical Data Analysis*, Wiley Series in Probability and Statistics, John Wiley & Sons, New York, NY, USA, 2nd edition, 2002. View at Zentralblatt MATH · View at MathSciNet - D. Collett,
*Modelling Binary Data*, Chapman & Hall/CRC Texts in Statistical Science Series, Chapman & Hall/CRC, Boca Raton, Fla, USA, 2nd edition, 2003. View at Zentralblatt MATH · View at MathSciNet - J. M. Landwehr, D. Pregibon, and A. C. Shoemaker, “Graphical methods for assessing logistic regression models,”
*Journal of the American Statisical Association*, vol. 79, no. 385, pp. 61–71, 1984. View at Publisher · View at Google Scholar · View at Zentralblatt MATH - SAS Institute,
*SAS/STAT User's Guide*, vol. 2, SAS Institute, Cary, NC, USA, 8th edition, 2000.