\SweaveOpts{engine=R,eps=FALSE,pdf=TRUE,width=8,strip.white=all}
\SweaveOpts{keep.source=TRUE}
\SweaveOpts{prefix=TRUE,prefix.string=figs/Multiple,include=TRUE}
\setkeys{Gin}{width=\textwidth}
<>=
options(lattice.theme = function() canonical.theme("pdf", color = FALSE),
str = strOptions(strict.width = "cut"), width=74,
show.signif.stars = FALSE)
library(splines)
library(lattice)
library(Matrix)
library(Rcpp)
library(minqa)
library(lme4)
fm03ML <- lmer(diameter ~ 1 + (1|plate) + (1|sample), Penicillin, REML=FALSE)
fm04 <- lmer(strength ~ 1 + (1|sample) + (1|batch), Pastes, REML=FALSE)
fm04a <- lmer(strength ~ 1 + (1|sample), Pastes, REML=FALSE)
if (file.exists("pr03.rda")) {
load("pr03.rda")
} else {
pr03 <- profile(fm03ML)
save(pr03, file = "pr03.rda")
}
if (file.exists("pr04.rda")) {load("pr04.rda")
} else {
pr04 <- profile(fm04)
save(pr04, file = "pr04.rda")
}
if (file.exists("pr04a.rda")) {load("pr04a.rda")
} else {
pr04a <- profile(fm04a)
save(pr04a, file = "pr04a.rda")
}
if (file.exists("fm05.rda")) {load("fm05.rda")
} else {
fm05 <- lmer(y ~ 1 + (1|s) + (1|d) + (1|dept:service), InstEval, REML=FALSE)
save(fm05, file = "fm05.rda")
}
if (file.exists("rr4.rda")) {load("rr4.rda")
} else {
rr4 <- ranef(fm05, condVar = TRUE, whichel = "dept:service")
save(rr4, file = "rr4.rda")
}
if (file.exists("fm05a.rda")) { load("fm05a.rda")
} else {
fm05a <- lmer(y ~ 1 + (1|s) + (1|d), InstEval, REML=FALSE)
save(fm05a, file = "fm05a.rda")
}
if (!file.exists("figs/Multiple-fm05Limage.png")) {
ttt <- getME(fm05, "L")
ttt@x[] <- 1
trellis.device(png, color = FALSE, file = "figs/Multiple-fm05Limage.png",
height=6, width=8, units='in', res=600)
print(image(ttt, sub=NULL, xlab=NULL, ylab=NULL, colorkey=FALSE))
rm(ttt)
dev.off()
}
@
\chapter{Models With Multiple Random-effects Terms}
\label{chap:Multiple}
The mixed models considered in the previous chapter had only one
random-effects term, which was a simple, scalar random-effects term,
and a single fixed-effects coefficient. Although such models can be
useful, it is with the facility to use multiple random-effects terms
and to use random-effects terms beyond a simple, scalar term that we
can begin to realize the flexibility and versatility of mixed models.
In this chapter we consider models with multiple simple, scalar
random-effects terms, showing examples where the grouping factors for
these terms are in completely crossed or nested or partially crossed
configurations. For ease of description we will refer to the random
effects as being crossed or nested although, strictly speaking, the
distinction between nested and non-nested refers to the grouping
factors, not the random effects.
\section{A Model With Crossed Random Effects}
\label{sec:crossedRE}
One of the areas in which the methods in the \package{lme4} package
for \R{} are particularly effective is in fitting models to
cross-classified data where several factors have random effects
associated with them. For example, in many experiments in psychology
the reaction of each of a group of subjects to each of a group of
stimuli or items is measured. If the subjects are considered to be a
sample from a population of subjects and the items are a sample from a
population of items, then it would make sense to associate random
effects with both these factors.
In the past it was difficult to fit mixed models with multiple,
crossed grouping factors to large, possibly unbalanced, data sets.
The methods in the \package{lme4} package are able to do this. To
introduce the methods let us first consider a small, balanced data set
with crossed grouping factors.
\subsection{The \texttt{Penicillin} Data}
\label{sec:Penicillin}
The \code{Penicillin} data are derived from Table~6.6, p.~144 of
\citet{davies72:_statis_method_in_resear_and_produc} where they are
described as coming from an investigation to
\begin{quote}
assess the variability between samples of penicillin by the
\emph{B.~subtilis} method. In this test method a bulk-innoculated
nutrient agar medium is poured into a Petri dish of approximately 90
mm. diameter, known as a plate. When the medium has set, six small
hollow cylinders or pots (about 4 mm. in diameter) are cemented onto
the surface at equally spaced intervals. A few drops of the
penicillin solutions to be compared are placed in the respective
cylinders, and the whole plate is placed in an incubator for a given
time. Penicillin diffuses from the pots into the agar, and this
produces a clear circular zone of inhibition of growth of the
organisms, which can be readily measured. The diameter of the zone
is related in a known way to the concentration of penicillin in the
solution.
\end{quote}
As with the \code{Dyestuff} data, we examine the structure
<>=
str(Penicillin)
@
and a summary
<>=
summary(Penicillin)
@
of the \code{Penicillin} data, then plot it
\begin{figure}[tbp]
\centering
<>=
print(dotplot(reorder(plate, diameter) ~ diameter, Penicillin, groups = sample,
ylab = "Plate", xlab = "Diameter of growth inhibition zone (mm)",
type = c("p", "a"), auto.key = list(columns = 3, lines = TRUE)))
@
\caption[Diameter of growth inhibition zone for 6 samples of
penicillin]{Diameter of the growth inhibition zone (mm) in the
\emph{B. subtilis} method of assessing the concentration of
penicillin. Each of 6 samples was applied to each of the 24 agar
plates. The lines join observations on the same sample.}
\label{fig:Penicillindot}
\end{figure}
(Fig.~\ref{fig:Penicillindot}).
The variation in the diameter is associated with the plates and with
the samples. Because each plate is used only for the six samples
shown here we are not interested in the contributions of specific
plates as much as we are interested in the variation due to plates and
in assessing the potency of the samples after accounting for this
variation. Thus, we will use random effects for the \code{plate}
factor. We will also use random effects for the \code{sample} factor
because, as in the dyestuff example, we are more interested in the
sample-to-sample variability in the penicillin samples than in the
potency of a particular sample.
In this experiment each sample is used on each plate. We say that the
\code{sample} and \code{plate} factors are \emph{crossed}, as opposed
to \emph{nested} factors, which we will describe in the next section.
By itself, the designation ``crossed'' just means that the factors are
not nested. If we wish to be more specific, we could describe these
factors as being \emph{completely crossed}, which means that we have at
least one observation for each combination of a level of \code{sample}
and a level of \code{plate}. We can see this in
Fig.~\ref{fig:Penicillindot} and, because there are moderate
numbers of levels in these factors, we can check it in a
cross-tabulation
<>=
xtabs(~ sample + plate, Penicillin)
@
Like the \code{Dyestuff} data, the factors in the \code{Penicillin}
data are balanced. That is, there are exactly the same number of
observations on each plate and for each sample and, furthermore, there
is the same number of observations on each combination of levels. In
this case there is exactly one observation for each combination of
sample and plate. We would describe the configuration of these two
factors as an unreplicated, completely balanced, crossed design.
In general, balance is a desirable but precarious property of a data
set. We may be able to impose balance in a designed experiment but we
typically cannot expect that data from an observation study will be
balanced. Also, as anyone who analyzes real data soon finds out,
expecting that balance in the design of an experiment will produce a
balanced data set is contrary to ``Murphy's Law''. That's why
statisticians allow for missing data. Even when we apply each of the
six samples to each of the 24 plates, something could go wrong for one
of the samples on one of the plates, leaving us without a measurement
for that combination of levels and thus an unbalanced data set.
\subsection{A Model For the \texttt{Penicillin} Data}
\label{sec:PenicillinModel}
A model incorporating random effects for both the \code{plate} and the
\code{sample} is straightforward to specify --- we include simple,
scalar random effects terms for both these factors.
<>=
(fm03 <- lmer(diameter ~ 1 + (1|plate) + (1|sample), Penicillin))
@
This model display indicates that the sample-to-sample variability has
the greatest contribution, then plate-to-plate variability and finally
the ``residual'' variability that cannot be attributed to either the
sample or the plate. These conclusions are consistent with what we
see in the \code{Penicillin} data plot
(Fig.~\ref{fig:Penicillindot}).
The prediction intervals on the random effects
(Fig.~\ref{fig:fm03ranef})
\begin{figure}[tbp]
\centering
<>=
qrr2 <- dotplot(ranef(fm03, condVar = TRUE), strip = FALSE)
print(qrr2[[1]], pos = c(0,0,1,0.75), more = TRUE)
print(qrr2[[2]], pos = c(0,0.65,1,1))
@
\caption[Random effects prediction intervals for model \code{fm03}]{95\%
prediction intervals on the random effects for model
\code{fm03} fit to the \code{Penicillin} data. The intervals in the
upper panel are those for levels of the \code{sample} factor. Those
in the lower panel correspond to levels of the \code{plate} factor.}
\label{fig:fm03ranef}
\end{figure}
confirm that the conditional distribution of the random effects for
\code{plate} has much less variability than does the conditional
distribution of the random effects for \code{sample}, in the sense that
the dots in the bottom panel have less variability than those in the
top panel. (Note the different horizontal axes for the two panels.)
However, the conditional distribution of the random effect for a
particular \code{sample}, say sample F, has less variability than the
conditional distribution of the random effect for a particular plate,
say plate m. That is, the lines in the bottom panel are wider than
the lines in the top panel, even after taking the different axis
scales into account. This is because the conditional distribution of
the random effect for a particular sample depends on 24 responses
while the conditional distribution of the random effect for a
particular plate depends on only 6 responses.
In chapter~\ref{chap:ExamLMM} we saw that a model with a single,
simple, scalar random-effects term generated a random-effects model
matrix, $\vec Z$, that is the matrix of indicators of the levels of
the grouping factor. When we have multiple, simple, scalar
random-effects terms, as in model \code{fm03}, each term generates a
matrix of indicator columns and these sets of indicators are
concatenated to form the model matrix $\vec Z$. The transpose of this
matrix, shown in Fig.~\ref{fig:fm03Ztimage}, contains rows of
indicators for each factor.
\begin{figure}[tbp]
\centering
<>=
print(image(getME(fm03, "Zt"), sub=NULL,xlab=NULL,ylab=NULL))
@
\caption[Image of the random-effects model matrix for
\texttt{fm03}]{Image of the transpose of the random-effects model
matrix, $\vec Z$, for model \code{fm03}. The non-zero elements,
which are all unity, are shown as darkened squares. The zero
elements are blank.}
\label{fig:fm03Ztimage}
\end{figure}
\begin{figure}[tbp]
\centering
<>=
print(image(getME(fm03, "Lambda"), sub=NULL,xlab=expression(Lambda),ylab=NULL),
split=c(1,1,3,1), more=TRUE)
print(image(tcrossprod(getME(fm03, "Zt")),sub=NULL,xlab="Z'Z",ylab=NULL),
split=c(2,1,3,1), more=TRUE)
print(image(getME(fm03, "L"), sub=NULL,xlab="L",ylab=NULL,colorkey=FALSE),
split=c(3,1,3,1))
@
\caption[Images of $\Lambda$, $\vec Z\trans\vec Z$ and $\vec L$ for
model \texttt{fm03}]{Images of the relative covariance factor,
$\Lambda$, the cross-product of the random-effects model matrix,
$\vec Z\trans\vec Z$, and the sparse Cholesky factor, $\vec L$, for
model \code{fm03}.}
\label{fig:fm03LambdaLimage}
\end{figure}
The relative covariance factor, $\Lambda_\theta$,
(Fig.~\ref{fig:fm03LambdaLimage}, left panel) is no longer a multiple
of the identity. It is now block diagonal, with two blocks, one of
size 24 and one of size 6, each of which is a multiple of the
identity. The diagonal elements of the two blocks are $\theta_1$ and
$\theta_2$, respectively. The numeric values of these parameters can
be obtained as
<>=
getME(fm03, "theta")
@
The first parameter is the relative standard deviation of the random
effects for \code{plate}, which has the value $0.84671/0.54992=1.53968$ at
convergence, and the second is the relative standard deviation of the
random effects for \code{sample} ($1.93157/0.54992=3.512443$).
Because $\Lambda_\theta$ is diagonal, the pattern of non-zeros in
$\Lambda_\theta\trans\vec Z\trans\vec Z\Lambda_\theta+\vec I$ will be
the same as that in $\vec Z\trans\vec Z$, shown in the middle panel of
Fig.~\ref{fig:fm03LambdaLimage}. The sparse Cholesky factor, $\vec L$,
shown in the right panel, is lower triangular and has non-zero elements
in the lower right hand corner in positions where $\vec Z\trans\vec Z$
has systematic zeros. We say that ``fill-in'' has occurred when
forming the sparse Cholesky decomposition. In this case there is a
relatively minor amount of fill but in other cases there can be a
substantial amount of fill and we shall take precautions so as to
reduce this, because fill-in adds to the computational effort in
determining the MLEs or the REML estimates.
A profile zeta plot (Fig.~\ref{fig:fm03prplot}) for the parameters in
model \code{fm03}
\begin{figure}[tbp]
\centering
<>=
print(xyplot(pr03, absVal=0, aspect=1.3, layout=c(4,1)))
@
\caption{Profile zeta plot of the parameters in model \code{fm03}.}
\label{fig:fm03prplot}
\end{figure}
leads to conclusions similar to those from Fig.~\ref{fig:fm1prof} for
model \code{fm01ML} in the previous chapter. The fixed-effect
parameter, $\beta_0$, for the \code{(Intercept)} term has symmetric
intervals and is over-dispersed relative to the normal distribution.
The logarithm of $\sigma$ has a good normal approximation but the
standard deviations of the random effects, $\sigma_1$ and $\sigma_2$,
are skewed. The skewness for $\sigma_2$ is worse than that for
$\sigma_1$, because the estimate of $\sigma_2$ is less precise than
that of $\sigma_1$, in both absolute and relative senses. For an
absolute comparison we compare the widths of the confidence intervals for
these parameters.
<>=
confint(pr03)
@
In a relative comparison we examine the ratio of the endpoints of
the interval divided by the estimate.
<>=
confint(pr03)[1:2,]/c(0.8455722, 1.770648)
@
(We have switched from the REML estimates shown in the display of
\code{fm03} to the ML estimates of the standard deviations.)
The lack of precision in the estimate of $\sigma_2$ is a consequence
of only having 6 distinct levels of the \code{sample} factor. The
\code{plate} factor, on the other hand, has 24 distinct levels. In
general it is more difficult to estimate a measure of spread, such as
the standard deviation, than to estimate a measure of location, such
as a mean, especially when the number of levels of the factor is
small. Six levels are about the minimum number required for obtaining
sensible estimates of standard deviations for simple, scalar random
effects terms.
The profile pairs plot (Fig.~\ref{fig:fm03prpairs}) shows
\begin{figure}[tbp]
\centering
<>=
print(splom(pr03))
@
\caption[Profile pairs plot of the parameters in model \code{fm03}.]
{Profile pairs plot for the parameters in model \code{fm03} fit to
the \code{Penicillin} data.}
\label{fig:fm03prpairs}
\end{figure}
patterns similar to those in Fig.~\ref{fig:fm1profpair} for pairs of
parameters in model \code{fm01} fit to the \code{Dyestuff} data. On
the $\zeta$ scale (panels below the diagonal) the profile traces are
nearly straight and orthogonal with the exception of the trace of
$\zeta(\sigma_2)$ on $\zeta(\beta_0)$ (the horizontal trace for the
panel in the $(4,2)$ position). The pattern of this trace is similar
to the pattern of the trace of $\zeta(\sigma_1)$ on $\zeta(\beta_0)$
in Fig.~\ref{fig:fm1profpair}. Moving $\beta_0$ from its estimate,
$\widehat{\beta}_0$, in either direction will increase the residual sum
of squares. The increase in the residual variability is reflected in
an increase of one or more of the dispersion parameters. The balanced
experimental design results in a fixed estimate of $\sigma$ and the
extra apparent variability must be incorporated into $\sigma_1$ or
$\sigma_2$.
Contours in panels of parameter pairs on the original scales
(i.e. panels above the diagonal) can show considerable distortion from
the ideal elliptical shape. For example, contours in the $\sigma_2$
versus $\sigma_1$ panel (the $(1,2)$ position) and the $\log(\sigma)$
versus $\sigma_2$ panel (in the $(2,3)$ position) are dramatically
non-elliptical. However, the distortion of the contours is not due to
these parameter estimates depending strongly on each other. It is almost
entirely due to the choice of scale for $\sigma_1$ and $\sigma_2$. When we
plot the contours on the scale of $\log(\sigma_1)$ and
$\log(\sigma_2)$ instead (Fig.~\ref{fig:lpr03pairs})
\begin{figure}[tbp]
\centering
<>=
print(splom(log(pr03)))
@
\caption[Profile pairs plot for model \code{fm03} (log scale)]
{Profile pairs plot for the parameters in model \code{fm03} fit to
the \code{Penicillin} data. In this plot the parameters
$\sigma_1$ and $\sigma_2$ are on the scale of the natural
logarithm, as is the parameter $\sigma$ in this and other profile
pairs plots.}
\label{fig:lpr03pairs}
\end{figure}
they are much closer to the elliptical pattern.
Conversely, if we tried to plot contours on the scale of $\sigma_1^2$
and $\sigma_2^2$ (not shown), they would be hideously distorted.
\section{A Model With Nested Random Effects}
\label{sec:NestedRE}
In this section we again consider a simple example, this time fitting
a model with \emph{nested} grouping factors for the random effects.
\subsection{The \texttt{Pastes} Data}
\label{sec:PastesData}
The third example from \citet[Table~6.5,
p.~138]{davies72:_statis_method_in_resear_and_produc} is described as
coming from
\begin{quote}
deliveries of a chemical paste product contained in casks where, in
addition to sampling and testing errors, there are variations in
quality between deliveries \dots As a routine, three casks selected
at random from each delivery were sampled and the samples were kept
for reference. \dots Ten of the delivery batches were sampled at
random and two analytical tests carried out on each of the 30
samples.
\end{quote}
The structure and summary of the \code{Pastes} data object are
<>=
str(Pastes)
summary(Pastes)
@
As stated in the description in
\begin{figure}[tbp]
\centering
<>=
print(image(xtabs(~ batch + sample, Pastes, sparse = TRUE),
sub = NULL, xlab = "sample", ylab = "batch"))
@
\caption[Cross-tabulation image of the \code{batch} and \code{sample}
factors]{Image of the cross-tabulation of the \code{batch} and
\code{sample} factors in the \code{Pastes} data.}
\label{fig:imagextabsPastes}
\end{figure}
\citet{davies72:_statis_method_in_resear_and_produc}, there are 30
samples, three from each of the 10 delivery batches. We have labelled
the levels of the \code{sample} factor with the label of the
\code{batch} factor followed by `a', `b' or `c' to distinguish the
three samples taken from that batch. The cross-tabulation produced by
the \code{xtabs} function, using the optional argument \code{sparse =
TRUE}, provides a concise display of the relationship.
<>=
xtabs(~ batch + sample, Pastes, drop = TRUE, sparse = TRUE)
@
Alternatively, we can use an image (Fig.~\ref{fig:imagextabsPastes})
of this cross-tabulation to visualize the structure.
When plotting the \code{strength} versus \code{batch} and
\code{sample} in the \code{Pastes} data we should remember that we
have two strength measurements on each of the 30 samples. It is
tempting to use the cask designation (`a', `b' and `c') to determine,
say, the plotting symbol within a \code{batch}. It would be fine
to do this within a batch but the plot would be misleading if we used
the same symbol for cask `a' in different batches. There is no
relationship between cask `a' in batch `A' and cask `a' in batch `B'.
The labels `a', `b' and `c' are used only to distinguish the three
samples within a batch; they do not have a meaning across batches.
\begin{figure}[tbp]
\centering
<>=
pp <- Pastes
pp <- within(pp, bb <- reorder(batch, strength))
pp <- within(pp, ss <- reorder(reorder(sample, strength),
as.numeric(batch)))
print(dotplot(ss ~ strength | bb, pp, pch = 21,
strip = FALSE, strip.left = TRUE, layout = c(1, 10),
scales = list(y = list(relation = "free")),
ylab = "Sample within batch", type = c("p", "a"),
xlab = "Paste strength", jitter.y = TRUE))
@
\caption[Strength of paste preparations by batch and sample]{Strength
of paste preparations according to the \code{batch} and the
\code{sample} within the batch. There were two strength
measurements on each of the 30 samples; three samples each from 10
batches.}
\label{fig:Pastesplot}
\end{figure}
In Fig.~\ref{fig:Pastesplot} we plot the two strength measurements
on each of the samples within each of the batches and join up the
average strength for each sample. The perceptive reader will have
noticed that the levels of the factors on the vertical axis in this
figure, and in Fig.~\ref{fig:Dyestuffdot} and
\ref{fig:Penicillindot}, have been reordered according to increasing
average response. In all these cases there is no inherent ordering of
the levels of the covariate such as \code{batch} or \code{plate}.
Rather than confuse our interpretation of the plot by determining the
vertical displacement of points according to a random ordering, we
impose an ordering according to increasing mean response. This allows
us to more easily check for structure in the data, including
undesirable characteristics like increasing variability of the
response with increasing mean level of the response.
In Fig.~\ref{fig:Pastesplot} we order the samples within each batch
separately, then order the batches according to increasing mean
strength.
Figure~\ref{fig:Pastesplot} shows considerable variability in strength
between samples relative to the variability within samples. There is
some indication of variability between batches, in addition to the
variability induced by the samples, but not a strong indication of a
batch effect. For example, batches I and D, with low mean strength
relative to the other batches, each contained one sample (I:b and D:c,
respectively) that had high mean strength relative to the other
samples. Also, batches H and C, with comparatively high mean batch
strength, contain samples H:a and C:a with comparatively low mean
sample strength. In Sect.~\ref{sec:TestingSig2is0} we will examine
the need for incorporating batch-to-batch variability, in addition to
sample-to-sample variability, in the statistical model.
\subsubsection{Nested Factors}
\label{sec:nestedcrossed}
Because each level of \code{sample} occurs with one and only one level
of \code{batch} we say that \code{sample} is \emph{nested within}
\code{batch}. Some presentations of mixed-effects models, especially
those related to \emph{multilevel modeling}~\citep{MLwiNUser:2000} or
\emph{hierarchical linear models}~\citep{Rauden:Bryk:2002}, leave the
impression that one can only define random effects with respect to
factors that are nested. This is the origin of the terms
``multilevel'', referring to multiple, nested levels of variability,
and ``hierarchical'', also invoking the concept of a hierarchy of
levels. To be fair, both those references do describe the use of
models with random effects associated with non-nested
factors, but such models tend to be treated as a special case.
The blurring of mixed-effects models with the concept of multiple,
hierarchical levels of variation results in an unwarranted emphasis on
``levels'' when defining a model and leads to considerable confusion.
It is perfectly legitimate to define models having random effects
associated with non-nested factors. The reasons for the emphasis on
defining random effects with respect to nested factors only are that
such cases do occur frequently in practice and that some of the
computational methods for estimating the parameters in the models can
only be easily applied to nested factors.
This is not the case for the methods used in the \package{lme4}
package. Indeed there is nothing special done for models with random
effects for nested factors. When random effects are associated with
multiple factors exactly the same computational methods are used
whether the factors form a nested sequence or are partially crossed or
are completely crossed.
%% FIXME: This is not true in lme4. Should this capability be added?
% A case of a nested sequence of ``grouping
% factors'' for the random effects (including the trivial case of only
% one such factor) is detected but this information does not change the
% course of the computation. It is available to be used as a diagnostic
% check. When the user knows that the grouping factors should be
% nested, she can check if they are indeed nested.
There is, however, one aspect of nested grouping factors that we
should emphasize, which is the possibility of a factor that is
\emph{implicitly nested} within another factor. Suppose, for example,
that the \code{sample} factor was defined as having three levels
instead of 30 with the implicit assumption that \code{sample} is
nested within \code{batch}. It may seem silly to try to distinguish
30 different batches with only three levels of a factor but,
unfortunately, data are frequently organized and presented like this,
especially in text books. The \code{cask} factor in the \code{Pastes}
data is exactly such an implicitly nested factor. If we
cross-tabulate \code{batch} and \code{cask}
<>=
xtabs(~ cask + batch, Pastes)
@
we get the impression that the \code{cask} and \code{batch} factors
are crossed, not nested. If we know that the cask should be
considered as nested within the batch then we should create a new
categorical variable giving the batch-cask combination, which is
exactly what the \code{sample} factor is. A simple way to create such
a factor is to use the interaction operator, `\code{:}', on the
factors. It is advisable, but not necessary, to apply \code{factor}
to the result thereby dropping unused levels of the interaction from
the set of all possible levels of the factor. (An ``unused level'' is
a combination that does not occur in the data.) A convenient code
idiom is
<>=
Pastes$sample <- with(Pastes, factor(batch:cask))
@ %$
or
<>=
Pastes <- within(Pastes, sample <- factor(batch:cask))
@
In a small data set like \code{Pastes} we can quickly detect a factor
being implicitly nested within another factor and take appropriate
action. In a large data set, perhaps hundreds of thousands of test
scores for students in thousands of schools from hundreds of school
districts, it is not always obvious if school identifiers are unique
across the entire data set or just within a district. If you are not
sure, the safest thing to do is to create the interaction factor, as
shown above, so you can be confident that levels of the
district:school interaction do indeed correspond to unique schools.
\subsection{Fitting a Model With Nested Random Effects}
\label{sec:fittingPastes}
Fitting a model with simple, scalar random effects for nested factors
is done in exactly the same way as fitting a model with random effects
for crossed grouping factors. We include random-effects terms for
each factor, as in
<>=
(fm04 <- lmer(strength ~ 1 + (1|sample) + (1|batch), Pastes, REML=FALSE))
@
Not only is the model specification similar for nested and crossed
\begin{figure}[tbp]
\centering
<>=
print(image(getME(fm04, "Lambda"), sub=NULL,xlab=expression(Lambda),ylab=NULL),
split=c(1,1,3,1), more=TRUE)
print(image(tcrossprod(getME(fm04, "Zt")),sub=NULL,xlab="Z'Z",ylab=NULL),
split=c(2,1,3,1), more=TRUE)
print(image(getME(fm04, "L"), sub=NULL,xlab="L",ylab=NULL,colorkey=FALSE),
split=c(3,1,3,1))
@
\caption[Images of $\Lambda$, $\vec Z\trans\vec Z$ and $\vec L$ for
model \texttt{fm04}]{Images of the relative covariance factor,
$\Lambda$, the cross-product of the random-effects model matrix,
$\vec Z\trans\vec Z$, and the sparse Cholesky factor, $\vec L$, for
model \code{fm04}.}
\label{fig:fm04LambdaLimage}
\end{figure}
factors, the internal calculations are performed according to the
methods described in Sect.~\ref{sec:definitions} for each model type.
Comparing the patterns in the matrices $\Lambda$, $\vec Z\trans\vec Z$
and $\vec L$ for this model (Fig.~\ref{fig:fm04LambdaLimage}) to those
in Fig.~\ref{fig:fm03LambdaLimage} shows that models with nested
factors produce simple repeated structures along the diagonal of the
sparse Cholesky factor, $\vec L$, after reordering the random effects
(we discuss this reordering later in Sect.~\ref{sec:fill-reducingP}).
This type of structure has the desirable property that there is no
``fill-in'' during calculation of the Cholesky factor. In other words,
the number of non-zeros in $\vec L$ is the same as the number of
non-zeros in the lower triangle of the matrix being factored,
$\Lambda\trans\vec Z\trans\vec Z\Lambda+\vec I$ (which, because
$\Lambda$ is diagonal, has the same structure as $\vec Z\trans\vec
Z$).
Fill-in of the Cholesky factor is not an important issue when we have
a few dozen random effects, as we do here. It is an important issue
when we have millions of random effects in complex configurations, as
has been the case in some of the models that have been fit using
\code{lmer}.
\subsection[Parameter Estimates for Model \code{fm04}]{Assessing
Parameter Estimates in Model \texttt{fm04}}
\label{sec:assessingfm04}
<>=
f3 <- "%.3f"
f4 <- "%.4f"
vc3 <- VarCorr(fm04)
stddev <- unname(c(sapply(vc3, function(el) attr(el, "stddev")),
attr(vc3, "sc")))
@
The parameter estimates are:
$\widehat{\sigma_1}=$\Sexpr{sprintf(f3,stddev[1])}, the standard
deviation of the random effects for \code{sample};
$\widehat{\sigma_2}=$\Sexpr{sprintf(f3,stddev[2])}, the standard
deviation of the random effects for \code{batch};
$\widehat{\sigma}=$\Sexpr{sprintf(f3,stddev[3])}, the standard
deviation of the residual noise term; and
$\widehat{\beta_0}=$\Sexpr{sprintf(f3,fixef(fm04))}, the overall mean
response, which is labeled \code{(Intercept)} in these models.
The estimated standard deviation for \code{sample} is nearly three
times as large as that for \code{batch}, which confirms what we saw in
Fig.~\ref{fig:Pastesplot}. Indeed our conclusion from
Fig.~\ref{fig:Pastesplot} was that there may not be a significant
batch-to-batch variability in addition to the sample-to-sample
variability.
Plots of the prediction intervals of the random effects
(Fig.~\ref{fig:fm04ranef})
\begin{figure}[tbp]
\centering
<>=
qrr3 <- dotplot(ranef(fm04, condVar = TRUE), strip = FALSE)
print(qrr3[[1]], pos = c(0,0,1,0.75), more = TRUE)
print(qrr3[[2]], pos = c(0,0.65,1,1))
@
\caption[Random effects prediction intervals for model \code{fm04}]{95\%
prediction intervals on the random effects for model
\code{fm04} fit to the \code{Pastes} data.}
\label{fig:fm04ranef}
\end{figure}
confirm this impression in that all the prediction intervals for the
random effects for \code{batch} contain zero.
\begin{figure}[tbp]
\centering
<>=
print(xyplot(pr04, absVal=0, aspect=1.3, layout=c(4,1)))
@
\caption{Profile zeta plots for the parameters in model \code{fm04}.}
\label{fig:fm04prplot}
\end{figure}
Furthermore, the profile zeta plot (Fig.~\ref{fig:fm04prplot})
shows that even the 50\% profile-based confidence interval on
$\sigma_2$ extends to zero.
Because there are several indications that $\sigma_2$ could reasonably
be zero, resulting in a simpler model incorporating random effects for
\code{batch} only, we perform a statistical test of this hypothesis.
\subsection{Testing $H_0:\sigma_2=0$ Versus $H_a:\sigma_2>0$}
\label{sec:TestingSig2is0}
One of the many famous statements attributed to Albert Einstein is
``Everything should be made as simple as possible, but not simpler.''
In statistical modeling this \emph{principal of parsimony} is embodied
in hypothesis tests comparing two models, one of which contains the
other as a special case. Typically, one or more of the parameters in
the more general model, which we call the \emph{alternative
hypothesis}, is constrained in some way, resulting in the restricted
model, which we call the \emph{null hypothesis}. Although we phrase
the hypothesis test in terms of the parameter restriction, it is
important to realize that we are comparing the quality of fits
obtained with two nested models. That is, we are not assessing
parameter values per se; we are comparing the model fit obtainable
with some constraints on parameter values to that without the
constraints.
Because the more general model, $H_a$, must provide a fit that is at
least as good as the restricted model, $H_0$, our purpose is to
determine whether the change in the quality of the fit is sufficient
to justify the greater complexity of model $H_a$. This comparison is
often reduced to a \emph{p-value}, which is the probability of seeing a
difference in the model fits as large as we did, or even larger, when,
in fact, $H_0$ is adequate. Like all probabilities, a p-value must be
between 0 and 1. When the p-value for a test is small (close to zero)
we prefer the more complex model, saying that we ``reject $H_0$ in
favor of $H_a$''. On the other hand, when the p-value is not small we
``fail to reject $H_0$'', arguing that there is a non-negligible
probability that the observed difference in the model fits could
reasonably be the result of random chance, not the inherent
superiority of the model $H_a$. Under these circumstances we prefer
the simpler model, $H_0$, according to the principal of parsimony.
These are the general principles of statistical hypothesis tests. To
perform a test in practice we must specify the criterion for comparing
the model fits, the method for calculating the p-value from an
observed value of the criterion, and the standard by which we will
determine if the p-value is ``small'' or not. The criterion is called
the \emph{test statistic}, the p-value is calculated from a
\emph{reference distribution} for the test statistic, and the standard
for small p-values is called the \emph{level} of the test.
In Sect.~\ref{sec:variability} we referred to likelihood ratio tests
(LRTs) for which the test statistic is the difference in the
deviance. That is, the LRT statistic is $d_0-d_a$ where $d_a$ is the
deviance in the more general ($H_a$) model fit and $d_0$ is the
deviance in the constrained ($H_0$) model. An approximate reference
distribution for an LRT statistic is the $\chi^2_\nu$ distribution
where $\nu$, the degrees of freedom, is determined by the number of
constraints imposed on the parameters of $H_a$ to produce $H_0$.
The restricted model fit
<>=
(fm04a <- lmer(strength ~ 1 + (1|sample), Pastes, REML=FALSE))
@
is compared to model \code{fm04} with the \code{anova} function
<>=
anova(fm04a, fm04)
@
which provides a p-value of
$\Sexpr{sprintf("%.4f",anova(fm04a,fm04)[["Pr(>Chisq)"]][2])}$.
Because typical
standards for ``small'' p-values are 5\% or 1\%, a p-value over 50\%
would not be considered significant at any reasonable level.
We do need to be cautious in quoting this p-value, however, because
the parameter value being tested, $\sigma_2=0$, is on the boundary of
set of possible values, $\sigma_2\ge 0$, for this parameter. The
argument for using a $\chi^2_1$ distribution to calculate a p-value
for the change in the deviance does not apply when the parameter value
being tested is on the boundary. As shown in
\citet[Sect.~2.5]{pinheiro00:_mixed_effec_model_in_s}, the p-value
from the $\chi^2_1$ distribution will be ``conservative'' in the sense
that it is larger than a simulation-based p-value would be. In the
worst-case scenario the $\chi^2$-based p-value will be twice as large
as it should be but, even if that were true, an effective p-value of
26\% would not cause us to reject $H_0$ in favor of $H_a$.
\subsection{Assessing the Reduced Model, \texttt{fm04a}}
\label{sec:assessReduced}
The profile zeta plots
\begin{figure}[tbp]
\centering
<>=
print(xyplot(pr04a, aspect=1.3, layout=c(3,1)))
@
\caption{Profile zeta plots for the parameters in model \code{fm04a}.}
\label{fig:fm04aprplot}
\end{figure}
for the remaining parameters in model \code{fm04a}
(Fig.~\ref{fig:fm04aprplot}) are similar to the corresponding panels in
Fig.~\ref{fig:fm04prplot}, as confirmed by the numerical values of the
confidence intervals.
<>=
confint(pr04)
confint(pr04a)
@
The confidence intervals on $\log(\sigma)$ and $\beta_0$ are similar
for the two models. The confidence interval on $\sigma_1$ is slightly
wider in model \code{fm04a} than in \code{fm04}, because the variability
that is attributed to \code{batch} in \code{fm04} is incorporated into
the variability due to \code{sample} in \code{fm04a}.
\begin{figure}[tbp]
\centering
<>=
print(splom(pr04a))
@
\caption[Profile pairs plot of the parameters in model \code{fm04a}.]
{Profile pairs plot for the parameters in model \code{fm04a} fit to
the \code{Pastes} data.}
\label{fig:fm04aprpairs}
\end{figure}
The patterns in the profile pairs plot (Fig.~\ref{fig:fm04aprpairs})
for the reduced model \code{fm04a} are similar to those in
Fig.~\ref{fig:fm1profpair}, the profile pairs plot for model
\code{fm01}.
\section{A Model With Partially Crossed Random Effects}
\label{sec:partially}
Especially in observational studies with multiple grouping factors,
the configuration of the factors frequently ends up neither nested nor
completely crossed. We describe such situations as having
\emph{partially crossed} grouping factors for the random effects.
Studies in education, in which test scores for students over time are
also associated with teachers and schools, usually result in partially
crossed grouping factors. If students with scores in multiple years
have different teachers for the different years, the student factor
cannot be nested within the teacher factor. Conversely, student and
teacher factors are not expected to be completely crossed. To have
complete crossing of the student and teacher factors it would be
necessary for each student to be observed with each teacher, which
would be unusual. A longitudinal study of thousands of students with
hundreds of different teachers inevitably ends up partially crossed.
In this section we consider an example with thousands of students and
instructors where the response is the student's evaluation of the
instructor's effectiveness. These data, like those from most large
observational studies, are quite unbalanced.
\subsection{The \texttt{InstEval} Data}
\label{sec:InstEval}
The \code{InstEval} data are from a special evaluation of lecturers by
students at the Swiss Federal Institute for Technology--Z\"{u}rich
(ETH--Z\"{u}rich), to determine who should receive the ``best-liked
professor'' award. These data have been slightly simplified and
identifying labels have been removed, so as to preserve anonymity.
The variables
<>=
## the default strict.width is "no" for back-compatibility
## instead, we could also *globally* set
## strOptions(strict.width = "cut")
##str(InstEval, strict.width = "cut")
str(InstEval)
@
have somewhat cryptic names. Factor \code{s} designates the student
and \code{d} the instructor. The \code{dept} factor is the department
for the course and \code{service} indicates whether the course was a
service course taught to students from other departments.
Although the response, \code{y}, is on a scale of 1 to 5,
<>=
xtabs(~ y, InstEval)
@
it is sufficiently diffuse to warrant treating it as if it were a
continuous response.
At this point we will fit models that have random effects for student,
instructor, and department (or the \code{dept:service} combination) to
these data. In the next chapter we will fit models incorporating
fixed-effects for instructor and department to these data.
<>=
(fm05 <- lmer(y ~ 1 + (1|s) + (1|d)+(1|dept:service), InstEval, REML=FALSE))
@
<>=
fm05
@
(Fitting this complex model to a moderately large data set takes less
than two minutes on a modest laptop computer purchased in 2006.
Although this is more time than required for earlier model fits, it is
a remarkably short time for fitting a model of this size and
complexity. In some ways it is remarkable that such a model can be
fit at all on such a computer.)
All three estimated standard deviations of the random effects are less
than $\widehat{\sigma}$, with $\widehat{\sigma}_3$, the estimated
standard deviation of the random effects for the \code{dept:service}
interaction, less than one-tenth the estimated residual standard
deviation.
\begin{figure}[tbp]
\centering
<>=
print(dotplot(rr4, strip = FALSE))
@
\caption[Random effects prediction intervals for model \code{fm05}]{95\%
prediction intervals on the random effects for the
\code{dept:service} factor in model \code{fm05} fit to the
\code{InstEval} data.}
\label{fig:fm05ranef}
\end{figure}
It is not surprising that zero is within all of the prediction
intervals on the random effects for this factor
(Fig.~\ref{fig:fm05ranef}). In fact, zero is close to the middle of
all these prediction intervals. However, the p-value for the LRT of
$H_0:\sigma_3=0$ versus $H_a:\sigma_3>0$
<>=
fm05a <- lmer(y ~ 1 + (1|s) + (1|d), InstEval, REML=0)
anova(fm05a,fm05)
@
<>=
anova(fm05a,fm05)
@
is highly significant. That is, we have very strong evidence that we
should reject $H_0$ in favor of $H_a$.
The seeming inconsistency of these conclusions is due to the large
sample size ($n=73421$). When a model is fit to a very large sample
even the most subtle of differences can be highly ``statistically
significant''. The researcher or data analyst must then decide if
these terms have practical significance, beyond the apparent
statistical significance.
The large sample size also helps to assure that the parameters have
good normal approximations. We could profile this model fit but doing
so would take a very long time and, in this particular case, the
analysts are more interested in a model that uses fixed-effects
parameters for the instructors, which we will describe in the next
chapter.
We could pursue other mixed-effects models here, such as using the
\code{dept} factor and not the \code{dept:service} interaction to
define random effects, but we will revisit these data in the next
chapter and follow up on some of these variations there.
\begin{figure}[tbp]
\centering
\includegraphics{figs/Multiple-fm05Limage}
\caption{Image of the sparse Cholesky factor, $\vec L$, from model \code{fm05}.}
\label{fig:fm05Limage}
\end{figure}
\subsection{Structure of $\vec L$ for model \texttt{fm05}}
\label{sec:fm05L}
Before leaving this model we examine the sparse Cholesky factor, $\vec
L$, (Fig.~\ref{fig:fm05Limage}), which is of size $4128\times4128$.
Even as a sparse matrix this factor requires a considerable amount of
memory,
<>=
object.size(getME(fm05, "L"))
unclass(round(object.size(getME(fm05, "L"))/2^20, 3)) # size in megabytes
@
but as a triangular dense matrix it would require nearly 10 times as
much. There are $(4128\times 4129)/2$ elements on and below the
diagonal, each of which would require 8 bytes of storage. A packed
lower triangular array would require
<>=
(8 * (4128 * 4129)/2)/2^20 # size in megabytes
@
megabytes. The more commonly used full rectangular storage requires
<>=
(8 * 4128^2)/2^20 # size in megabytes
@
megabytes of storage.
The number of nonzero elements in this matrix that must be updated for
each evaluation of the deviance is
<>=
nnzero(as(getME(fm05, "L"), "sparseMatrix"))
@
Comparing this to \Sexpr{(4128*4129)/2}, the number of elements that
must be updated in a dense Cholesky factor, we can see why the sparse
Cholesky factor provides a much more efficient evaluation of the
profiled deviance function.
\section{Chapter Summary}
\label{sec:MultSummary}
A simple, scalar random effects term in an \code{lmer} model formula
is of the form \code{(1|fac)}, where \code{fac} is an expression whose
value is the \emph{grouping factor} of the set of random effects
generated by this term. Typically, \code{fac} is simply the name of a
factor, such as in the terms \code{(1|sample)} or \code{(1|plate)} in
the examples in this chapter. However, the grouping factor can be the
value of an expression, such as \code{(1|dept:service)} in the last
example.
Because simple, scalar random-effects terms can differ only in the
description of the grouping factor we refer to configurations such as
crossed or nested as applying to the terms or to the random effects,
although it is more accurate to refer to the configuration as applying
to the grouping factors.
A model formula can include several such random effects terms. Because
configurations such as nested or crossed or partially crossed grouping
factors are a property of the data, the specification in the model
formula does not depend on the configuration. We simply include
multiple random effects terms in the formula specifying the model.
One apparent exception to this rule occurs with implicitly nested
factors, in which the levels of one factor are only meaningful within
a particular level of the other factor. In the \code{Pastes}
data, levels of the \code{cask} factor are only meaningful within a
particular level of the \code{batch} factor. A model formula of
<>=
strength ~ 1 + (1|cask) + (1|batch)
@
would result in a fitted model that did not appropriately reflect the
sources of variability in the data. Following the simple rule that
the factor should be defined so that distinct experimental or
observational units correspond to distinct levels of the factor will
avoid such ambiguity.
For convenience, a model with multiple, nested random-effects terms
can be specified as
<>=
strength ~ 1 + (1|batch/cask)
@
which internally is re-expressed as
<>=
strength ~ 1 + (1|batch) + (1|batch:cask)
@
We will avoid terms of the form \code{(1|batch/cask)}, preferring
instead an explicit specification with simple, scalar terms based on
unambiguous grouping factors.
The \code{InstEval} data, described in Sec.~\ref{sec:InstEval},
illustrate some of the characteristics of the real data to which
mixed-effects models are now fit. There is a large number of
observations associated with several grouping factors; two of which,
student and instructor, have a large number of levels and are
partially crossed. Such data are common in sociological and
educational studies but until now it has been very difficult to fit
models that appropriately reflect such a structure. Much of the
literature on mixed-effects models leaves the impression that multiple
random effects terms can only be associated with nested grouping
factors. The resulting emphasis on hierarchical or multilevel
configurations is an artifact of the computational methods used to fit
the models, not the models themselves.
The parameters of the models fit to small data sets have properties
similar to those for the models in the previous chapter. That is,
profile-based confidence intervals on the fixed-effects parameter,
$\beta_0$, are symmetric about the estimate but overdispersed
relative to those that would be calculated from a normal distribution and
the logarithm of the residual standard deviation, $\log(\sigma)$, has
a good normal approximation. Profile-based confidence intervals for
the standard deviations of random effects ($\sigma_1$, $\sigma_2$,
etc.) are symmetric on a logarithmic scale except for those that could
be zero.
Another observation from the last example is that, for data sets with
a very large numbers of observations, a term in a model may be
``statistically significant'' even when its practical significance is
questionable.
\section*{Exercises}
\addcontentsline{toc}{section}{Exercises}
These exercises use data sets from the \package{MEMSS} package for
\R{}. Recall that to access a particular data set, you must either attach
the package
<>=
library(MEMSS)
@
or load just the one data set
<>=
data(ergoStool, package = "MEMSS")
@
We begin with exercises using the \code{ergoStool} data from the
\package{MEMSS} package. The analysis and graphics in these exercises
is performed in Chap.~\ref{chap:Covariates}. The purpose of these
exercises is to see if you can use the material from this chapter to
anticipate the results quoted in the next chapter.
\begin{prob}\label{pr:2ergoexamine}
Check the documentation, the structure (\code{str}) and a summary of
the \code{ergoStool} data from the \package{MEMSS} package. (If you
are familiar with the Star Trek television series and movies, you may
want to speculate about what, exactly, the ``Borg scale'' is.) Use
<>=
xtabs(~ Type + Subject, ergoStool)
@
to determine if these factors are nested, partially crossed or
completely crossed. Is this a replicated or an unreplicated design?
\end{prob}
\begin{prob}\label{pr:2ergoplot} %% Give hints on this question.
Create a plot, similar to Fig.~\ref{fig:Penicillindot}, showing the
effort by subject with lines connecting points corresponding to the
same stool types. Order the levels of the \code{Subject} factor by
increasing average \code{effort}.
\end{prob}
\begin{prob}\label{pr:2ergofitre}
The experimenters are interested in comparing these specific stool
types. In the next chapter we will fit a model with fixed-effects
for the \code{Type} factor and random effects for \code{Subject},
allowing us to perform comparisons of these specific types. At
this point fit a model with random effects for both \code{Type} and
\code{Subject}. What are the relative sizes of the estimates of the
standard deviations, $\widehat{\sigma}_1$ (for \code{Subject}),
$\widehat{\sigma}_2$ (for \code{Type}) and $\widehat{\sigma}$ (for
the residual variability)?
\end{prob}
\begin{prob}
Refit the model using maximum likelihood. Check the parameter
estimates and, in the case of the fixed-effects parameter,
$\beta_0$, its standard error. In what ways have the parameter
estimates changed? Which parameter estimates have not changed?
\end{prob}
\begin{prob}
Profile the fitted model and construct 95\% profile-based confidence
intervals on the parameters. (Note that you will get the same
profile object whether you start with the REML fit or the ML fit.
There is a slight advantage in starting with the ML fit.) Is the
confidence interval on $\sigma_1$ close to being symmetric about its
estimate? Is the confidence interval on $\sigma_2$ close to being
symmetric about its estimate? Is the corresponding interval on
$\log(\sigma_1)$ close to being symmetric about its estimate?
\end{prob}
\begin{prob}
Create the profile zeta plot for this model. For which parameters
are there good normal approximations?
\end{prob}
\begin{prob}
Create a profile pairs plot for this model. Comment on the shapes
of the profile traces in the transformed ($\zeta$) scale and the
shapes of the contours in the original scales of the parameters.
\end{prob}
\begin{prob}
Create a plot of the 95\% prediction intervals on the random effects
for \code{Type} using
<>=
dotplot(ranef(fm, which = "Type", condVar = TRUE), aspect = 0.2,
strip = FALSE)
@
(Substitute the name of your fitted model for \code{fm} in the call to
\code{ranef}.) Is there a clear winner among the stool types?
(Assume that lower numbers on the Borg scale correspond to less
effort).
\end{prob}
\begin{prob}
Create a plot of the 95\% prediction intervals on the random effects
for \code{Subject}.
\end{prob}
\begin{prob}
Check the documentation, the structure (\code{str}) and a summary of
the \code{Meat} data from the \package{MEMSS} package. Use a
cross-tabulation to discover whether \code{Pair} and \code{Block}
are nested, partially crossed or completely crossed.
\end{prob}
\begin{prob}
Use a cross-tabulation
<>=
xtabs(~ Pair + Storage, Meat)
@
to determine whether \code{Pair} and \code{Storage} are nested,
partially crossed or completely crossed.
\end{prob}
\begin{prob}
Fit a model of the \code{score} in the \code{Meat} data with random
effects for \code{Pair}, \code{Storage} and \code{Block}.
\end{prob}
\begin{prob}
Plot the prediction intervals for each of the three sets of random effects.
\end{prob}
\begin{prob}
Profile the parameters in this model. Create a profile zeta plot.
Does including the random effect for \code{Block} appear to be
warranted. Does your conclusion from the profile zeta plot agree
with your conclusion from examining the prediction intervals for the
random effects for \code{Block}?
\end{prob}
\begin{prob}
Refit the model without random effects for \code{Block}. Perform a
likelihood ratio test of $H_0:\sigma_3=0$ versus $H_a:\sigma_3>0$.
Would you reject $H_0$ in favor of $H_a$ or fail to reject $H_0$?
Would you reach the same conclusion if you adjusted the p-value for
the test by halving it, to take into account the fact that 0 is on
the boundary of the parameter region?
\end{prob}
\begin{prob}
Profile the reduced model (i.e. the one without random effects for
\code{Block}) and create profile zeta and profile pairs plots. Can
you explain the apparent interaction between $\log(\sigma)$ and
$\sigma_1$? (This is a difficult question.)
\end{prob}