In this section, we present a very general procedure which allows a
linear hypothesis to be tested, i.e.,
a linear restriction, either on a vector mean
or on the coefficient
of a linear model.
The presented technique covers
many of the practical testing problems on means or regression coefficients.
Linear hypotheses are of the form
with known matrices
and
with
.
EXAMPLE 7.7
Let
![$\mu=(\mu_{1},\mu_{2})^{\top}$](mvahtmlimg2133.gif)
.
The hypothesis that
![$\mu_{1}=\mu_{2}$](mvahtmlimg1078.gif)
can be equivalently written as:
The general idea is to test a normal population
(restricted model) against the
full model
where no restrictions are put on
.
Due to the properties of the
multinormal, we can easily adapt the Test Problems 1 and 2 to this new
situation. Indeed we know, from Theorem 5.2, that
, where
and
.
Testing the null
, is the same as testing
. The appropriate statistics are
and
which
can be derived from the original statistics
and
available
from
:
Here the difference between the sample mean and the tested value is
.
We are now in the situation to proceed to Test Problem 5 and 6.
By (7.2) we have that, under
:
and we reject
if this test statistic is too large at the desired
significance level.
EXAMPLE 7.8
We consider hypotheses on partitioned mean vectors
![$\mu
=\left({\mu_{1}\atop\mu_{2}}\right)$](mvahtmlimg2146.gif)
. Let us
first look at
for
![$N_{2p}({\mu_1 \choose \mu_2},\left({\Sigma\atop
0}{0\atop\Sigma}\right))$](mvahtmlimg2148.gif)
with known
![$\Sigma$](mvahtmlimg869.gif)
. This is
equivalent to
![$\data{A}=(\data{I},-\data{I})$](mvahtmlimg2149.gif)
,
![$a=(0,\dots,0)^{\top}\in\mathbb{R}^p$](mvahtmlimg2150.gif)
and leads to:
Another example is the test whether
![$\mu_1=0$](mvahtmlimg2152.gif)
, i.e.,
for
![$N_{2p}({\mu_1 \choose \mu_2},\left({\Sigma\atop
0}{0\atop\Sigma}\right))$](mvahtmlimg2148.gif)
with known
![$\Sigma$](mvahtmlimg869.gif)
.
This is equivalent to
![$\data{A}\mu =a$](mvahtmlimg2130.gif)
with
![$\data{A}=(\data{I},0)$](mvahtmlimg2154.gif)
,
and
![$a=(0,\dots,0)^{\top}\in\mathbb{R}^p$](mvahtmlimg2150.gif)
.
Hence:
From Corollary (5.4) and under
it follows immediately that
![$\displaystyle (n-1)(\data{A}\overline x-a)^{\top}
(\data{ASA}^{\top})^{-1}(\data{A}\overline x-a)
\sim$](mvahtmlimg2158.gif) |
![$\textstyle T^2(q,n-1)$](mvahtmlimg2159.gif) |
|
(7.9) |
since indeed under
,
is independent of
EXAMPLE 7.9
Let's come back again to the bank data set and suppose
that we want to test if
![$\mu_4 =\mu_5$](mvahtmlimg2162.gif)
, i.e., the hypothesis that
the lower border mean equals the larger border mean for the forged bills.
In this case:
The test statistic is:
The observed value is
![$13.638$](mvahtmlimg2165.gif)
which is significant at the
![$5\%$](mvahtmlimg747.gif)
level.
In many situations,
independent sampling units are observed at
different times
or under
different experimental conditions (different treatments,...).
So here we repeat
one-dimensional measurements on
different
subjects. For instance, we observe the results from
students
taking
different exams. We end up with a
matrix.
We can thus consider the situation where
we have
i.i.d. from a normal distribution
when there are
repeated measurements.
The hypothesis of interest in this case is that
there are no treatment effects,
.
This hypothesis is a direct application of Test Problem 6.
Indeed, introducing an appropriate matrix transform on
we have:
![\begin{displaymath}
H_0:\ {\cal{C}}\mu=0\ \textrm{where}\ {\cal{C}} ((p-1)\times...
... \vdots &\vdots \\
0 & \cdots & 0 &1 &-1
\end{array}\right).
\end{displaymath}](mvahtmlimg2169.gif) |
(7.10) |
Note that in many cases one of the experimental conditions is the ``control'' (a placebo,
standard drug or reference condition). Suppose it is the first component.
In that case one is interested in studying
differences to the control variable.
The matrix
has therefore a different form
By (7.9) the null hypothesis will be rejected if:
As a matter of fact,
is the mean of the random
variable
Simultaneous confidence intervals for linear combinations of the mean of
have been derived above in (7.7).
For all
, with probability
we have:
Due to the nature of the problem here, the row sums of the elements in
are zero:
, therefore
is a vector
whose sum of elements vanishes. This is called a
contrast. Let
.
We have
.
The result above thus provides for all contrasts of
, and
simultaneous confidence intervals at level
Examples of contrasts for
are
or
or even
when the control is to be compared with the
mean of 3 different treatments.
EXAMPLE 7.10
Bock (1975) considers the evolution of the vocabulary of children
from the eighth through eleventh grade.
The data set contains the scores of a vocabulary test of 40 randomly chosen
children that are observed from grades 8 to 11.
This is a repeated measurement situation,
![$(n=40, p=4)$](mvahtmlimg2188.gif)
,
since the same children were observed from grades 8 to 11.
The statistics of interest are:
Suppose we are interested in the yearly evolution of the children.
Then the matrix
![${\cal{C}}$](mvahtmlimg2170.gif)
providing successive differences of
![$\mu_j$](mvahtmlimg2190.gif)
is:
The value of the test statistic is
![$F_{\textrm{obs}}=53.134$](mvahtmlimg2192.gif)
which is highly significant for
![$F_{3.37}.$](mvahtmlimg2193.gif)
There are significant differences between the successive means.
However, the analysis of the contrasts shows the following simultaneous
![$95\%$](mvahtmlimg2073.gif)
confidence intervals
Thus, the rejection of
![$H_0$](mvahtmlimg918.gif)
is mainly due to the difference between the
childrens' performances in the first and second year.
The confidence intervals for the following contrasts may also be of interest:
They show that
![$\mu_1$](mvahtmlimg2196.gif)
is different from the average of the 3 other years
(the same being true for
![$\mu_4$](mvahtmlimg2197.gif)
)
and
![$\mu_4$](mvahtmlimg2197.gif)
turns out to be higher than
![$\mu_2$](mvahtmlimg2198.gif)
(and of course higher than
![$\mu_1$](mvahtmlimg2196.gif)
).
Test Problem 7 illustrates how the likelihood ratio
can be applied
when testing a linear restriction on
the coefficient
of a linear model.
It is also shown how a transformation of the test statistic leads to
an exact
test as presented in Chapter 3.
The constrained maximum likelihood estimators under
are
(Exercise 3.24):
for
and
.
The estimate
denotes the unconstrained MLE as before.
Hence, the LR statistic is
where
is the number of elements of
. This problem
also has an exact
-test since
EXAMPLE 7.11
Let us continue with the ``classic blue'' pullovers. We can once more
test if
![$\beta=0$](mvahtmlimg835.gif)
in the regression of sales on prices. It holds that
The LR statistic here is
which is not significant for the
![$\chi_{1}^2$](mvahtmlimg2208.gif)
distribution. The
![$F$](mvahtmlimg64.gif)
-test statistic
is also not significant. Hence, we can assume independence of
sales and prices (alone). Recall that this conclusion has to be revised if we
consider the prices together with advertisement costs and hours of
sales managers.
Recall the different conclusion that was made in Example 7.6
when we rejected
and
. The rejection there came from
the fact that the pair of values
was rejected. Indeed, if
the estimator of
would be
and this is too far from
.
EXAMPLE 7.12
Let us now consider the multivariate regression in the ``classic blue''
pullovers example. From Example
3.15 we know that the
estimated parameters in the model
are
Hence, we could postulate the approximate relation:
which means in practice that augmenting the price by 20 EUR
requires the advertisement costs to increase by 10 EUR in order to keep
the number of pullovers sold constant. Vice versa, reducing the price by
20 EUR yields the same result as before if we reduced the advertisement
costs by 10 EUR. Let us now test whether the hypothesis
is valid. This is equivalent to
The LR statistic in this case is equal to (
MVAlrtest.xpl
)
the
![$F$](mvahtmlimg64.gif)
statistic is
Hence, in both cases we will not reject the null hypothesis.
In many situations, we want to compare two groups of individuals for whom a set
of
characteristics has been observed. We have two random samples
and
from two distinct
-variate normal populations. Several testing issues can be addressed in this
framework. In Test Problem 8
we will first test the hypothesis of equal mean
vectors in the two groups
under the assumption of equality of the two covariance
matrices. This task can be solved by adapting Test Problem 2.
In Test Problem 9 a procedure for testing the equality of the
two covariance matrices is presented. If the covariance matrices differ,
the procedure of Test Problem 8 is no longer valid.
If the equality of the covariance matrices is rejected,
an easy rule for comparing two means with no
restrictions on the covariance matrices
is provided in Test Problem 10.
TEST PROBLEM 8
Assume that
![$X_{i1} \sim N_{p}(\mu_{1},\Sigma)$](mvahtmlimg2222.gif) , with
![$ i=1,\cdots,n_{1}$](mvahtmlimg2223.gif) and
![$X_{j2} \sim N_p(\mu_{2},\Sigma)$](mvahtmlimg2224.gif) , with
![$ j=1,\cdots,n_{2}$](mvahtmlimg2225.gif) , where all the
variables are independent.
|
Both samples provide the statistics
and
,
.
Let
. We have
![\begin{displaymath}
(\bar{x}_{1}-\bar{x}_{2})\sim N_{p}\left(\delta, \frac{n_1+n_2}{n_{1}n_{2}}\Sigma\right)
\end{displaymath}](mvahtmlimg2231.gif) |
(7.11) |
![\begin{displaymath}
n_{1}S_{1}+n_{2}S_{2}\sim W_{p}(\Sigma, n_{1}+n_{2}-2).
\end{displaymath}](mvahtmlimg2232.gif) |
(7.12) |
Let
=
be the weighted
mean of
and
. Since the two samples are
independent and since
is independent of
(for
) it follows that
is independent of
Hence, Theorem 5.8 applies
and leads to a
-distribution:
![\begin{displaymath}
\frac{n_{1}n_{2}(n_{1}+n_{2}-2)}{(n_{1}+n_{2})^2}
\left\{ ( ...
..._{1}-\bar{x}_{2} ) -\delta\right\})
\sim T^2(p, n_{1}+n_{2}-2)
\end{displaymath}](mvahtmlimg2237.gif) |
(7.13) |
or
This result, as in Test Problem 2, can be used to test
:
=0 or to construct a confidence region for
.
The rejection region is given by:
![\begin{displaymath}
\frac{n_{1}n_{2}(n_{1}+n_{2}-p-1)}{p(n_{1}+n_{2})^2}
\left(\...
...r{x}_{1}
-\bar{x}_{2}\right) \ge F_{1-\alpha ;p,n_1+n_2-p-1}.
\end{displaymath}](mvahtmlimg2241.gif) |
(7.14) |
A
confidence region for
is given by the ellipsoid
centered at
and the simultaneous confidence intervals for all linear combinations
of the elements of
are given by
In particular we have at the
level, for
,
![\begin{displaymath}
\delta_j \in(\bar{x}_{1j}-\bar{x}_{2j}) \pm \sqrt{\frac{p(n...
...2}
-p-1)(n_{1}n_{2})} F_{1-\alpha ;p,n_{1}+n_{2}-p-1} s_{jj}}.
\end{displaymath}](mvahtmlimg2246.gif) |
(7.15) |
EXAMPLE 7.13
Let us come back to the questions raised in Example
7.5.
We compare the means of assets (
![$X_1$](mvahtmlimg14.gif)
)
and of sales (
![$X_2$](mvahtmlimg13.gif)
) for two sectors,
energy (group 1) and manufacturing (group 2).
With
![$n_1=15$](mvahtmlimg2247.gif)
,
![$n_2=10$](mvahtmlimg2248.gif)
, and
![$p=2$](mvahtmlimg298.gif)
we obtain the statistics:
and
so that
The observed value of the test statistic (
7.14) is
![$F=2.7036$](mvahtmlimg2252.gif)
.
Since
![$F_{0.95;2,22}=3.4434$](mvahtmlimg2253.gif)
the hypothesis of equal means of the two groups is
not rejected although it would be rejected at a less severe level
(
![$F>F_{0.90;2,22}=2.5613$](mvahtmlimg2254.gif)
).
The 95% simultaneous confidence intervals for the differences
(
MVAsimcidif.xpl
) are given by
EXAMPLE 7.14
In order to illustrate the presented test procedures it is interesting to
analyze some simulated data.
This simulation will point out the importantce of the covariances in
testing means.
We created 2 independent normal
samples in
![$\mathbb{R}^4$](mvahtmlimg2256.gif)
of sizes
![$n_1=30$](mvahtmlimg2257.gif)
and
![$n_2=20$](mvahtmlimg2258.gif)
with:
One may consider this as an example of
![${X}=({X}_1,\ldots ,{X}_n)^{\top}$](mvahtmlimg2261.gif)
being the students' scores from 4 tests, where the 2 groups
of students were subjected to two different methods of teaching.
First we simulate the two samples with
![$\Sigma={\data I}_4$](mvahtmlimg2262.gif)
and
obtain the statistics:
The test statistic (
7.14) takes the value
![$F=60.65$](mvahtmlimg2264.gif)
which
is highly significant: the small variance allows the difference
to be detected even
with these relatively moderate sample sizes.
We conclude (at the 95% level) that:
which confirms that the means for
![${X}_1$](mvahtmlimg2266.gif)
and
![${X}_4$](mvahtmlimg2267.gif)
are different.
Consider now a different simulation scenario where the standard deviations are 4
times larger:
.
Here we obtain:
Now the test statistic takes the value 1.54 which is no longer significant
(
![$F_{0.95,4,45} =2.58$](mvahtmlimg2270.gif)
).
Now we cannot reject the null hypothesis (which we know to be false!) since
the increase in variances prohibits the detection of differences
of such magnitude.
The following situation illustrates once more the role of the covariances
between covariates.
Suppose that
as above but with
(this corresponds to a negative correlation
).
We have:
The value of
![$F$](mvahtmlimg64.gif)
is
![$3.853$](mvahtmlimg2274.gif)
which is significant at the 5% level
(p-value = 0.0089). So the null hypothesis
![$\delta=\mu_1-\mu_2=0$](mvahtmlimg2275.gif)
is outside
the 95% confidence ellipsoid. However, the simultaneous confidence intervals, which
do not take the covariances into account
are given by:
They contain the null value (see Remark
7.1 above) although they are very
asymmetric for
![$\delta_1$](mvahtmlimg2277.gif)
and
![$\delta_4$](mvahtmlimg2278.gif)
.
EXAMPLE 7.15
Let us compare the vectors of means of the forged and the genuine bank notes.
The matrices
![$\data{S}_f$](mvahtmlimg2032.gif)
and
![$\data{S}_g$](mvahtmlimg2279.gif)
were given in Example
3.1 and since here
![$n_f=n_g=100$](mvahtmlimg2280.gif)
,
![$\data{S}$](mvahtmlimg687.gif)
is the simple average of
![$\data{S}_f$](mvahtmlimg2032.gif)
and
![$\data{S}_g: \data{S}=\frac{1}{2}\left(\data{S}_f+\data{S}_g\right)$](mvahtmlimg2281.gif)
.
The test statistic is given by (
7.14) and turns out to be
![$F=391.92$](mvahtmlimg2283.gif)
which is
highly significant for
![$F_{6,193}$](mvahtmlimg2284.gif)
. The 95% simultaneous confidence intervals for the
differences
![$\delta_j=\mu_{gj}-\mu_{fj},\ j=1,\ldots ,p$](mvahtmlimg2285.gif)
are:
All of the components (except for the first one)
show significant differences in the means. The main
effects are taken by the lower border
![$(X_4)$](mvahtmlimg1702.gif)
and the diagonal
![$(X_6)$](mvahtmlimg2287.gif)
.
The preceding test implicitly uses the fact that the two samples are
extracted from two different populations with common variance
.
In this case, the test statistic (7.14) measures the distance between
the two centers of gravity of the two groups w.r.t. the common metric given by
the pooled variance matrix
. If
no such matrix exists.
There are no satisfactory test procedures for testing the equality of variance matrices
which are robust with respect to normality assumptions of the populations.
The following test extends Bartlett's test for equality of variances in the
univariate case. But this test is known to be very sensitive to
departures from normality.
TEST PROBLEM 9
(Comparison of Covariance Matrices)
Let
![$X_{ih} \sim N_{p}(\mu_{h},\Sigma_{h})$](mvahtmlimg2289.gif) ,
![$i=1,\dots,n_{h}$](mvahtmlimg2290.gif) , ![$h=1,\dots,k$](mvahtmlimg2291.gif)
be independent random variables,
|
Each subsample provides
, an estimator of
, with
Under
,
(Section 5.2), where
is the common covariance matrix
and
.
Let
be the weighted
average of the
(this is in fact the MLE of
when
is true). The likelihood ratio test leads to the statistic
![\begin{displaymath}
-2\log\lambda = n\log\mid S\mid-\sum_{h=1}^{k}n_{h}\log\mid S_{h}\mid
\end{displaymath}](mvahtmlimg2299.gif) |
(7.16) |
which under
is approximately distributed as a
where
.
EXAMPLE 7.16
Let's come back to Example
7.13,
where the mean of assets and sales have been compared
for companies from the energy and manufacturing sector
assuming that
![$\Sigma_1 =\Sigma_2$](mvahtmlimg2302.gif)
. The test of
![$\Sigma_1 =\Sigma_2$](mvahtmlimg2302.gif)
leads to the value of the test statistic
![\begin{displaymath}
-2\log\lambda=0.9076
\end{displaymath}](mvahtmlimg2303.gif) |
(7.17) |
which is not significant (p-value for a
![$\chi^2_3 =0.82$](mvahtmlimg2304.gif)
). We cannot reject
![$H_0$](mvahtmlimg918.gif)
and the
comparison of the means performed above is valid.
EXAMPLE 7.17
Let us compare the covariance matrices of the
forged and the genuine bank notes (the matrices
![$S_f$](mvahtmlimg2305.gif)
and
![$S_g$](mvahtmlimg2306.gif)
are shown in Example
3.1).
A first look seems to suggest that
![$\Sigma_1 \neq \Sigma_2$](mvahtmlimg2307.gif)
.
The pooled variance
![$S$](mvahtmlimg2080.gif)
is given by
![$\data{S}=\frac{1}{2}\left(\data{S}_f+\data{S}_g\right)$](mvahtmlimg2308.gif)
since here
![$n_f=n_g$](mvahtmlimg2309.gif)
.
The test statistic here is
![$-2\log\lambda =127.21$](mvahtmlimg2310.gif)
,
which is highly significant
![$\chi^2$](mvahtmlimg1720.gif)
with 21 degrees of freedom. As expected,
we reject the hypothesis of equal covariance matrices,
and as a result the procedure for comparing the two means
in Example
7.15 is not valid.
What can we do with unequal covariance matrices? When both
and
are large, we
have a simple solution:
TEST PROBLEM 10
(Comparison of two means, unequal covariance matrices, large samples)
Assume that
![${X}_{i1} \sim N_{p}(\mu_{1},\Sigma_1)$](mvahtmlimg2313.gif) , with
![$ i=1,\cdots,n_{1}$](mvahtmlimg2223.gif) and
![${X}_{j2} \sim N_p(\mu_{2},\Sigma_2)$](mvahtmlimg2314.gif) , with
![$ j=1,\cdots,n_{2}$](mvahtmlimg2225.gif) are independent random variables.
|
Letting
, we have
Therefore, by (5.4)
Since
is a consistent estimator of
for
, we have
![\begin{displaymath}
(\bar{x}_1-\bar{x}_2)^{\top}\left(\frac{\data{S}_1}{n_1}+\fr...
...
(\bar{x}_1-\bar{x}_2) \stackrel{{\mathcal{L}}}{\to} \chi^2_p.
\end{displaymath}](mvahtmlimg2321.gif) |
(7.18) |
This can be used in place of (7.13) for testing
, defining a confidence
region for
or constructing simultaneous confidence intervals for
.
For instance, the rejection region at the level
will be
![\begin{displaymath}
(\bar{x}_1-\bar{x}_2)^{\top}\left(\frac{\data{S}_1}{n_1}+\fr...
...}{n_2}\right)^{-1}
(\bar{x}_1-\bar{x}_2)> \chi^2_{1-\alpha ;p}
\end{displaymath}](mvahtmlimg2323.gif) |
(7.19) |
and the
simultaneous confidence intervals for
,
are:
![\begin{displaymath}
\delta_j \in (\bar{x}_1-\bar{x}_2) \pm \sqrt{\chi^2_{1-\alph...
...left(\frac{s^{(1)}_{jj}}{n_1}+\frac{s^{(2)}_{jj}}{n_2}\right)}
\end{displaymath}](mvahtmlimg2325.gif) |
(7.20) |
where
is the
element of the matrix
.
This may be compared to (7.15) where the pooled variance was used.
REMARK 7.2
We see, by comparing the statistics (
7.19) with (
7.14),
that we measure here
the distance between
![$\bar{x}_1$](mvahtmlimg2328.gif)
and
![$\bar{x}_2$](mvahtmlimg2329.gif)
using the metric
![$\left(\frac{\data{S}_1}{n_1}+
\frac{\data{S}_2}{n_2}\right)$](mvahtmlimg2330.gif)
. It should be noticed that when
![$n_1 = n_2$](mvahtmlimg2331.gif)
, the two methods are
essentially the same since then
![$\data{S}=\frac{1}{2}\left(\data{S}_1+\data{S}_2\right)$](mvahtmlimg2332.gif)
.
If the covariances are different but have the same eigenvectors
(different eigenvalues),
one can apply the common principal component (CPC) technique,
see Chapter
9.
EXAMPLE 7.18
Let us use the last test to compare the forged and the genuine bank notes again (
![$n_1$](mvahtmlimg2311.gif)
and
![$n_2$](mvahtmlimg2312.gif)
are both large). The test statistic (
7.19) turns out to be 2436.8 which is again
highly significant. The 95% simultaneous confidence intervals are:
showing that all the components except the first are different from zero, the larger difference
coming from
![$X_6$](mvahtmlimg104.gif)
(length of the diagonal) and
![$X_4$](mvahtmlimg11.gif)
(lower border). The results are very
similar to those obtained in Example (
7.15). This is due to the fact that here
![$n_1 = n_2$](mvahtmlimg2331.gif)
as we already mentioned in the remark above.
Profile Analysis
Another useful application of Test Problem 6
is the repeated measurements problem applied to two independent groups.
This problem arises in practice when
we observe repeated measurements of characteristics (or measures of the same type under different
experimental conditions) on the different groups which have to be compared.
It is important that the
measures (the ``profile'') are comparable and
in particular are reported in the same units.
For instance, they may be measures of blood pressure at
different
points in time, one group being the
control group and the other the group receiving a new treatment.
The observations may be the
scores obtained from
different tests of two different experimental groups.
One is then interested in comparing the profiles of each group:
the profile being just the vectors
of the means of the
responses (the comparison may be visualized in a two dimensional graph
using the parallel coordinate plot introduced in Section 1.7).
We are thus in the same statistical situation
as for the comparison of two means:
where all variables are independent.
Suppose the two population profiles look like Figure 7.1.
The following questions are of interest:
- Are the profiles similar in the sense of being parallel (which means no interaction
between the treatments and the groups)?
- If the profiles are parallel, are they at the same level?
- If the profiles are parallel, is there any treatment effect, i.e.,
are the profiles horizontal?
The above questions are easily translated into
linear constraints on the means and a test statistic can be obtained
accordingly.
Parallel Profiles
Let
be a
matrix defined as
The hypothesis to be tested is
From (7.11), (7.12) and Corollary 5.4
we know that under
:
![\begin{displaymath}
\frac{n_1n_2}{(n_1+n_2)^2}(n_1+n_2-2)\left\{{\cal{C}}(\bar{x...
...})^{-1}{\cal{C}}
(\bar{x}_1-\bar{x}_2) \sim T^2(p-1,n_1+n_2-2)
\end{displaymath}](mvahtmlimg2340.gif) |
(7.21) |
where
is the pooled covariance matrix.
The hypothesis is rejected if
The question of equality of the two levels is meaningful only if the two profiles are parallel.
In the case of interactions (rejection of
), the two populations react differently to the
treatments and the question of the level has no meaning.
The equality of the two levels can be formalized as
since
and
Using Corollary 5.4 we have that:
The rejection region is
If it is rejected that the profiles are parallel, then two independent
analyses should be
done on the two groups using the repeated measurement approach.
But if it is accepted that they are parallel,
then we can exploit the information
contained in both groups (eventually at different levels)
to test a treatment effect, i.e., if the two profiles are horizontal.
This may be written as:
Consider the average profile
:
Clearly,
Now it is not hard to prove that
with
implies that
So under parallel, horizontal profiles we have
From Corollary 5.4 we again obtain
![\begin{displaymath}
(n_1+n_2-2)({\cal{C}}\bar{x})^{\top}({\cal{C}}\data{S}{\cal{C}}^{\top})^{-1}{\cal{C}}\bar{x}\sim T^2(p-1, n_1+n_2-2).
\end{displaymath}](mvahtmlimg2357.gif) |
(7.23) |
This leads to the rejection region of
, namely
EXAMPLE 7.19
Morrison (1990) proposed a test in which the results of
4 sub-tests of the Wechsler Adult Intelligence Scale
(WAIS) are compared for 2 categories of people:
group 1 contains
![$n_1=37$](mvahtmlimg2359.gif)
people who do not have a senile factor and
group 2 contains
![$n_2=12$](mvahtmlimg2360.gif)
people who have a senile factor.
The four WAIS sub-tests are
![$X_1$](mvahtmlimg14.gif)
(information),
![$X_2$](mvahtmlimg13.gif)
(similarities),
![$X_3$](mvahtmlimg221.gif)
(arithmetic) and
![$X_4$](mvahtmlimg11.gif)
(picture completion).
The relevant statistics are
The test statistic for testing if the two profiles are parallel is
![$F=0.4634$](mvahtmlimg2362.gif)
, which is not significant (
![$p$](mvahtmlimg34.gif)
-value
![$=0.71$](mvahtmlimg2363.gif)
).
Thus it is accepted that the two are parallel.
The second test statistic
(testing the equality of the levels of the 2 profiles)
is
![$F =17.21$](mvahtmlimg2364.gif)
, which is highly
significant (
![$p$](mvahtmlimg34.gif)
-value
![$\approx 10^{-4}$](mvahtmlimg2365.gif)
).
The global level of the test for the non-senile
people is superior to the senile group.
The final test (testing the horizontality of the average
profile) has the test statistic
![$F= 53.32$](mvahtmlimg2366.gif)
,
which is also highly significant (
![$p$](mvahtmlimg34.gif)
-value
![$ \approx 10^{-14}$](mvahtmlimg2367.gif)
).
This implies that there are substantial differences among the means of the
different subtests.
Summary
![$\ast$](mvahtmlimg108.gif)
-
Hypotheses about
can often be written as
, with
matrix
, and vector
.
![$\ast$](mvahtmlimg108.gif)
-
The hypothesis
for
with
known leads to
, where
is the number of elements in
.
![$\ast$](mvahtmlimg108.gif)
-
The hypothesis
for
with
unknown leads to
, where
is the number of elements in
and we have an exact test
![$\ast$](mvahtmlimg108.gif)
-
The hypothesis
for
with
unknown leads to
, with
being the length of
and with