Two-Way Tables (Reading: Faraway (2006), section 4.1~4.2)

   

Two-by-Two Tables.

The data shown below were collected as part of a quality improvement study at a semiconductor factory.

Quality No Particles Particles Total
Good 320 14 334
Bad 80 36 116
Total 400 50 450

A sample of wafers was drawn and cross-classified according to whether a particle was found on the die that produced the wafer and whether the wafer was good or bad. First, let us set up the data in a convenient form for analysis:

> y <- c(320,14,80,36)
> particle <- gl(2,1,4,labels=c("no","yes"))
> quality <- gl(2,2,labels=c("good","bad"))
> wafer <- data.frame(y,particle,quality)
> wafer

    y particle quality

1 320       no    good

2  14      yes    good

3  80       no     bad

4  36      yes     bad

We will need the data in this form with one observation per line for our model fitting, but usually we prefer to observe its table form:

> (ov <- xtabs(y ~ quality+particle))

       particle

quality  no yes

   good 320  14

   bad   80  36

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The data might have arisen under several possible sampling schemes:

  1. observe the manufacturing process for a certain period of time and observed 450 wafers Þ all the numbers are random Þ we could use a Poisson model

  2. decide to sample 450 wafers Þ 450 is a fixed number and other numbers are random Þ we could use a multinomial model

  3. select 400 wafers without particles and 50 wafers with particles Þ 400, 50, and 450 are fixed numbers and other numbers are random Þ we could use a binomial model

  4. select 400 wafers without particles and 50 wafers with particles that also included, by design, 334 good wafers and 116 bad ones Þ 400, 50, 334, 116, and 450 are fixed numbers and other numbers are random Þ we could use a hypergeometric model

The main question of interest concerning these data is whether the presence of particles on the wafer affects the quality outcome. We shall see in the following that all four sampling schemes lead to exactly the same conclusion.


Sampling Scheme 1: Poisson Model.
Under sampling scheme 1, it would be natural to view these outcomes, Yij, 1£i£I, 1£j£J, occurring at different rates, lpij,  where l is an unknown value of a size variable and pij is the probability that a randomly selected subject falls in the ijth cell. We can form a Poisson model for these rates, i.e., Yij~Poisson(lpij). Suppose that we fit a Poisson GLM with only main effect terms:

> modl <- glm(y ~ particle+quality, poisson)
> summary(modl)

Call:

glm(formula = y ~ particle + quality, family = poisson)

 

Deviance Residuals:

     1       2       3       4 

 1.324  -4.350  -2.370   5.266 

 

Coefficients:

            Estimate Std. Error z value Pr(>|z|)   

(Intercept)   5.6934     0.0572  99.535   <2e-16 ***

particleyes  -2.0794     0.1500 -13.863   <2e-16 ***

qualitybad   -1.0576     0.1078  -9.813   <2e-16 ***

---

Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

 

(Dispersion parameter for poisson family taken to be 1)

 

    Null deviance: 474.10  on 3  degrees of freedom

Residual deviance:  54.03  on 1  degrees of freedom

AIC: 83.774

 

Number of Fisher Scoring iterations: 5

Notice that:

 

Sampling Scheme 2: Multinomial Model.
Suppose that we assume that the total sample size was fixed at 450 and that the frequency of the four possible outcomes was recorded. In this circumstances, it is natural to use a multinomial distribution to model the response. There is a close connection between multinomial and Poisson models as follows. Let Yij be the independent Poisson random variables with means lpij, 1£i£I, 1£j£J, then the joint distribution of (Y11, Y12, ..., YIJ|SijYij=y++) is multinomial(y++, p11, p12, ..., pIJ). This essentially means that we can model multinomial GLM's using a Poisson GLM. The inferences for pij in the multinomial model would coincide with that in the Poisson GLM. For example, let us compute the MLE of pi+ and p+j under the multinomial model, which are SjYij/y++ and SiYij/y++, respectively:  

> (pp <- prop.table( xtabs(y ~ particle)))

particle

       no       yes

0.8888889 0.1111111

> (qp <- prop.table( xtabs(y ~ quality)))

quality

     good       bad

0.7422222 0.2577778

and the fitted values, Yij-hat, of each cells under the independence hypothesis, i.e., pij=pi+p+j, are y++(pi+-hat)(p+j-hat):

> (fv <- outer(qp,pp)*450)

       particle

quality       no      yes

   good 296.8889 37.11111

   bad  103.1111 12.88889

Notice that the fitted values are the same as that in the Poisson GLM with only main effect terms.

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To test whether the fitted values are close to the observed values, we can compare this model against the saturated model, for which the MLE of cell counts are the observed values Yij. So, the log-likelihood-ratio test statistic for independence is 2SiSjYijlog[Yij/(Yij-hat)] [Note: it is because the log-likelihood of multinomial is SiSjYijlog(pij) ], which computes to:

> 2*sum(ov*log(ov/fv))

[1] 54.03045

This is the same deviance we observed in the Poisson model with only main effect terms. So, we see that the test for independence in the multinomial model coincides with the test for no interaction in the Poisson model. The latter test is easier to execute in R, so we shall usually take that approach.

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One alternative to the deviance is the Pearson X2 statistic SiSj[(Oij-Eij)2/Eij]:

> sum( (ov-fv)^2/fv)

[1] 62.81231

which is also the same as that in the Poisson model with only main effect terms. We can apply Yates' continuity correction on the Pearson X2 statistic, which gives superior results for small samples. This correction is implemented in:

> prop.test(ov)

        2-sample test for equality of proportions with continuity correction

 

data:  ov

X-squared = 60.1239, df = 1, p-value = 8.907e-15

alternative hypothesis: two.sided

95 percent confidence interval:

 0.1757321 0.3611252

sample estimates:

   prop 1    prop 2

0.9580838 0.6896552

The X2 statistic with continuity correction is 60.1239 for one degree of freedom, which clearly rejects the null, i.e., independence hypothesis.

 

Sampling Scheme 3: Binomial Model.

Under the sampling scheme 3, it would be natural to view the presence of the particle as a cause that affects the quality of wafer. In other words, we would view the variable quality as the response and the variable particle as a predictor. When the number of wafers with no particles is fixed at 400 and the number with particles at 50, we can use a binomial model for the response for both groups. To fit a binomial GLM to the response in R, we need to form a two-column matrix with the 1st column representing the number of "successes" and the 2nd column the number of "failures":

> (m <- matrix(y,nrow=2))

     [,1] [,2]

[1,]  320   80

[2,]   14   36

Let us fit a binomial GLM that only contains a constant term, which suggests that the probability of the response is the same in both the particle and no particle group, i.e., a homogeneous model:

> modb <- glm(m ~ 1, family=binomial)

> summary(modb)

Call:

glm(formula = m ~ 1, family = binomial)

 

Deviance Residuals:

     1       2 

 2.715  -6.831 

 

Coefficients:

            Estimate Std. Error z value Pr(>|z|)   

(Intercept)   1.0576     0.1078   9.813   <2e-16 ***

---

Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

 

(Dispersion parameter for binomial family taken to be 1)

 

    Null deviance: 54.03  on 1  degrees of freedom

Residual deviance: 54.03  on 1  degrees of freedom

AIC: 66.191

 

Number of Fisher Scoring iterations: 3

Notice that:

Because either row or column marginal totals are fixed in this sampling scheme, the marginal totals in the table does not carry any useful information about the marginal proportions pi+ and p+j. Therefore, it does not make sense to test equivalence of marginal proportions for data obtained from the sampling scheme with fixed marginal totals.

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Sampling Scheme 4: Hypergeometric Model.

Under the sampling scheme 4, if we fix any number in the table, say Y11, the remaining three numbers are completely determined because the row and column marginal totals are known. In the circumstance, the distribution of Y11 under the null hypothesis: pij=pi+p+j is a hypergeometric distribution. Because there is only a limited number of values which Y11 can possibly take, we can compute the total probability of all outcomes that are "more extreme" than the one observed. The total probability is then the p-value. The method is called Fisher's exact test

> fisher.test(ov)

        Fisher's Exact Test for Count Data

 

data:  ov

p-value = 2.955e-13

alternative hypothesis: true odds ratio is not equal to 1

95 percent confidence interval:

  5.090628 21.544071

sample estimates:

odds ratio

  10.21331

Notice that:


Larger Two-Way Table.
Snee (1974) presents the data on 592 students cross-classified by hair and eye color. Let us read the data into R and take a look of it:

> haireye <- read.table("haireye.txt")
> haireye

     y   eye  hair

1    5 green BLACK

2   29 green BROWN

3   14 green   RED

4   16 green BLOND

5   15 hazel BLACK

6   54 hazel BROWN

7   14 hazel   RED

8   10 hazel BLOND

9   20  blue BLACK

10  84  blue BROWN

11  17  blue   RED

12  94  blue BLOND

13  68 brown BLACK

14 119 brown BROWN

15  26 brown   RED

16   7 brown BLOND

The data is more conveniently displayed using:

> (ct <- xtabs(y ~ hair + eye, haireye))

       eye

hair    blue brown green hazel

  BLACK   20    68     5    15

  BLOND   94     7    16    10

  BROWN   84   119    29    54

  RED     17    26    14    14

(Q: what information have you found by merely looking at the table?)

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We can execute the usual Pearson's X2 test for independence as:

> summary(ct)

Call: xtabs(formula = y ~ hair + eye, data = haireye)

Number of cases in table: 592

Number of factors: 2

Test for independence of all factors:

        Chisq = 138.29, df = 9, p-value = 2.325e-25

where

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Now, let us fit a Poisson GLM containing only main effect terms:

> modc <- glm(y ~ hair + eye, family=poisson, haireye)
> summary(modc)

Call:

glm(formula = y ~ hair + eye, family = poisson, data = haireye)

 

Deviance Residuals:

    Min       1Q   Median       3Q      Max 

-7.3263  -2.0646  -0.2120   1.2353   6.1724 

 

Coefficients:

            Estimate Std. Error z value Pr(>|z|)   

(Intercept)  3.66926    0.11055  33.191  < 2e-16 ***

hairBLOND    0.16206    0.13089   1.238  0.21569   

hairBROWN    0.97386    0.11294   8.623  < 2e-16 ***

hairRED     -0.41945    0.15279  -2.745  0.00604 **

eyebrown     0.02299    0.09590   0.240  0.81054   

eyegreen    -1.21175    0.14239  -8.510  < 2e-16 ***

eyehazel    -0.83804    0.12411  -6.752 1.46e-11 ***

---

Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

 

(Dispersion parameter for poisson family taken to be 1)

 

    Null deviance: 453.31  on 15  degrees of freedom

Residual deviance: 146.44  on  9  degrees of freedom

AIC: 241.04

 

Number of Fisher Scoring iterations: 5

Notice that:

One option for displaying contingency table data is a graphical tool, dotchart:

> dotchart(ct)

An alternative graphical tool is the mosaic plot, which divides the plot region according to the frequency of each level in a recursive manner:

> mosaicplot(ct,color=T,main="")

In the plot,

(Q: what patterns will you see in the dotchart and mosaic plots if the two variables are independent? what if they have equivalent marginal proportions? Generate some tables with the two properties and draw the plots.)

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