With Repeated Measures Part 1

This is a two part document. For the second part go to Mixed-Models-for-Repeated-Measures2.html

When we have a design in which we have both random and fixed variables, we have what is often called a mixed model. Mixed models have begun to play an important role in statistical analysis and offer many advantages over more traditional analyses. At the same time they are more complex and the syntax for software analysis is not always easy to set up. I will break this paper up into two papers because there are a number of designs and design issues to consider. This document will deal with the use of what are called mixed models (or linear mixed models, or hierarchical linear models, or many other things) for the analysis of what we normally think of as a simple repeated measures analysis of variance. Future documents will deal with mixed models to handle single-subject design (particularly multiple baseline designs) and nested designs.

A large portion of this document has benefited from Chapter 15 in Maxwell & Delaney (2004) *Designing Experiments and Analyzing Data.* They have one of the clearest discussions that I know. I am going a step beyond their example by including a between-groups factor as well as a within-subjects (repeated measures) factor. For now my purpose is to show the relationship between mixed models and the analysis of variance. The relationship is far from perfect, but it gives us a known place to start. More importantly, it allows us to see what we gain and what we lose by going to mixed models. In some ways I am going through the Maxwell & Delaney chapter backwards, because I am going to focus primarily on the use of the repeated command in SAS **Proc Mixed**. I am doing that because it fits better with the transition from ANOVA to mixed models.

My motivation for this document came from a question asked by Rikard Wicksell at Karolinska University in Sweden. He had a randomized clinical trial with two treatment groups and measurements at pre, post, 3 months, and 6 months. His problem is that some of his data were missing. He considered a wide range of possible solutions, including "last trial carried forward," mean substitution, and listwise deletion. In some ways listwise deletion appealed most, but it would mean the loss of too much data. One of the nice things about mixed models is that we can use all of the data we have. If a score is missing, it is just missing. It has no effect on other scores from that same patient.

Another advantage of mixed models is that we don't have to be consistent about time. For example, and it does not apply in this particular example, if one subject had a follow-up test at 4 months while another had their follow-up test at 6 months, we simply enter 4 (or 6) as the time of follow-up. We don't have to worry that they couldn't be tested at the same intervals.

A third advantage of these models is that we do not have to assume sphericity or compound symmetry in the model. We can do so if we want, but we can also allow the model to select its own set of covariances or use covariance patterns that we supply. I will start by assuming sphericity because I want to show the parallels between the output from mixed models and the output from a standard repeated measures analysis of variance. I will then delete a few scores and show what effect that has on the analysis. I will compare the standard analysis of variance model with a mixed model. Finally I will use Expectation Maximization (EM) and Multiple Imputation (MI) to impute missing values and then feed the newly complete data back into a repeated measures ANOVA to see how those results compare. (If you want to read about those procedures, I have a web page on them at Missing.html).

I have created data to have a number of characteristics. There are two groups - a Control group and a Treatment group, measured at 4 times. These times are labeled as 1 (pretest), 2 (one month posttest), 3 (3 months follow-up), and 4 (6 months follow-up). I created the treatment group to show a sharp drop at post-test and then sustain that drop (with slight regression) at 3 and 6 months. The Control group declines slowly over the 4 intervals but does not reach the low level of the Treatment group. There are noticeable individual differences in the Control group, and some subjects show a steeper slope than others. In the Treatment group there are individual differences in level but the slopes are not all that much different from one another. You might think of this as a study of depression, where the dependent variable is a depression score (e.g. Beck Depression Inventory) and the treatment is drug versus no drug. If the drug worked about as well for all subjects the slopes would be comparable and negative across time. For the control group we would expect some subjects to get better on their own and some to stay depressed, which would lead to differences in slope for that group. These facts are important because when we get to the random coefficient mixed model the individual differences will show up as variances in intercept, and any slope differences will show up as a significant variance in the slopes. For the standard ANOVA, and for mixed models using the **Repeated** command, the differences in level show up as a Subject effect and we assume that the slopes are comparable across subjects.

The program and data used below are available at the following links. I explain below the differences between the data files.

Wicksell.sas WicksellWide.dat WicksellLong.dat WicksellWideMiss.dat WicksellLongMiss.dat

Many of the printouts that follow were generated using SAS **Proc Mixed**, but I give the SPSS commands as well. (I also give syntax for R, but I warn you that running this problem under R, even if you have Pinheiro & Bates (2000), is very difficult. I only give these commands for one analysis, but they can be modified for related analyses.

The data follow. Notice that to set this up for ANOVA (**Proc GLM**) we read in the data one subject at a time. (You can see this in the data shown. Files like this are often said to be in "wide format" for reasons that are easy to see.) This will become important because we will not do that for mixed models. There we will use the "long" format.

Group Subj Time0 Time1 Time3 Time6 1 1 296 175 187 192 1 2 376 329 236 76 1 3 309 238 150 123 1 4 222 60 82 85 1 5 150 271 250 216 1 6 316 291 238 144 1 7 321 364 270 308 1 8 447 402 294 216 1 9 220 70 95 87 1 10 375 335 334 79 1 11 310 300 253 140 1 12 310 245 200 120 2 13 282 186 225 134 2 14 317 31 85 120 2 15 362 104 144 114 2 16 338 132 91 77 2 17 263 94 141 142 2 18 138 38 16 95 2 19 329 62 62 6 2 20 292 139 104 184 2 21 275 94 135 137 2 22 150 48 20 85 2 23 319 68 67 12 2 24 300 138 114 174

A plot of the data follows:

The cell means and standard errors follow:

----------------------------------------- group=Control --------------------------------- Variable N Mean Std Dev Minimum Maximum -------------------------------------------------------------------------------- time1 12 304.3333333 79.0642240 150.0000000 447.0000000 time2 12 256.6666667 107.8503452 60.0000000 402.0000000 time3 12 215.7500000 76.5044562 82.0000000 334.0000000 time4 12 148.8333333 71.2866599 76.0000000 308.0000000 xbar 12 231.3958333 67.9581638 112.2500000 339.7500000 --------------------------------------------------------------------------------- ---------------------------------------- group=Treatment ---------------------------------- Variable N Mean Std Dev Minimum Maximum --------------------------------------------------------------------------------- time1 12 280.4166667 69.6112038 138.0000000 362.0000000 time2 12 94.5000000 47.5652662 31.0000000 186.0000000 time3 12 100.3333333 57.9754389 16.0000000 225.0000000 time4 12 106.6666667 55.7939934 6.0000000 184.0000000 xbar 12 145.4791667 42.9036259 71.7500000 206.7500000 ----------------------------------------------------------------------------------

The results of a standard repeated measures analysis of variance with no missing data and using SAS **Proc GLM** follow. You would obtain the same results using the SPSS Univariate procedure. Because I will ask for a polynomial trend analysis, I have told it to recode the levels as 0, 1, 3, 6 instead of 1, 2, 3, 4. I did not need to do this, but it seemed truer to the experimental design. It does not affect the standard summary table. (I give the entire data entry parts of the program here, but will leave it out in future code.)

Options nodate nonumber nocenter formdlim = '-'; libname lib 'C:\Users\Dave\Documents\Webs\StatPages\More_Stuff\MixedModelsRepeated'; Title 'Analysis of Wicksell complete data'; data lib.WicksellWide; infile 'C:\Users\Dave\Documents\Webs\StatPages\More_Stuff\MixedModelsRepeated\WicksellWide.dat' firstobs = 1; input group subj time1 time2 time3 time4; xbar = (time1+time2+time3+time4)/4; run; Proc Format; Value group 1 = 'Control' 2 = 'Treatment' ; run; Proc Means data = lib.WicksellWide; Format group group.; var time1 -- time4 xbar; by group; run; Title 'Proc GLM with Complete Data'; proc GLM ; class group; model time1 time2 time3 time4 = group/ nouni; repeated time 4 (0 1 3 6) polynomial /summary printm; run; --------------------------------------------------------------------------------------- Proc GLM with Complete Data The GLM Procedure Repeated Measures Analysis of Variance Tests of Hypotheses for Between Subjects Effects Source DF Type III SS Mean Square F Value Pr > F group 1 177160.1667 177160.1667 13.71 0.0012 Error 22 284197.4583 12918.0663 ---------------------------------------------------------------------------------------- Proc GLM with Complete Data The GLM Procedure Repeated Measures Analysis of Variance Univariate Tests of Hypotheses for Within Subject Effects Adj Pr > F Source DF Type III SS Mean Square F Value Pr > F G - G H - F time 3 373802.7083 124600.9028 45.14 <.0001 <.0001 <.0001 time*group 3 74654.2500 24884.7500 9.01 <.0001 0.0003 0.0001 Error(time) 66 182201.0417 2760.6218 Greenhouse-Geisser Epsilon 0.7297 Huynh-Feldt Epsilon 0.8503 ---------------------------------------------------------------------------------------- Proc GLM with Complete Data Analysis of Variance of Contrast Variables time_N represents the nth degree polynomial contrast for time Contrast Variable: time_1 Source DF Type III SS Mean Square F Value Pr > F Mean 1 250491.4603 250491.4603 54.27 <.0001 group 1 2730.0179 2730.0179 0.59 0.4500 Error 22 101545.1885 4615.6904 Contrast Variable: time_2 Source DF Type III SS Mean Square F Value Pr > F Mean 1 69488.21645 69488.21645 35.37 <.0001 group 1 42468.55032 42468.55032 21.62 0.0001 Error 22 43224.50595 1964.75027 Contrast Variable: time_3 Source DF Type III SS Mean Square F Value Pr > F Mean 1 53823.03157 53823.03157 31.63 <.0001 group 1 29455.68182 29455.68182 17.31 0.0004 Error

Here we see that each of the effects in the overall analysis is significant. We don't care very much about the group effect because we expected both groups to start off equal at pre-test. What is important is the interaction, and it is significant at *p* = .0001. Clearly the drug treatment is having a differential effect on the two groups, which is what we wanted to see. The fact that the Control group seems to be dropping in the number of symptoms over time is to be expected and not exciting, although we could look at these simple effects if we wanted to. We would just run two analyses, one on each group. I would not suggest pooling the variances to calculate *F*, though that would be possible.

In the printout above I have included tests on linear, quadratic, and cubic trend that will be important later. However you have to read this differently than you might otherwise expect. The first test for the linear component shows an *F* of 54.27 for "mean" and an *F* of 0.59 for "group." Any other software that I have used would replace "mean" with "Time" and "group" with "Group × Time." In other words we have a significant linear trend over time, but the linear × group contrast is not significant. I don't know why they label them that way. (Well, I guess I do, but it's not the way that I would do it.) I should also note that my syntax specified the intervals for time, so that SAS is not assuming equally spaced intervals. The fact that the linear trend was not significant for the interaction means that both groups are showing about the same linear trend. But notice that there is a significant interaction for the quadratic.

The use of mixed models represents a substantial difference from the traditional analysis of variance. For balanced designs (which roughly translates to equal cell sizes) the results will come out to be the same, assuming that we set the analysis up appropriately. But the actual statistical approach is quite different and ANOVA and mixed models will lead to different results whenever the data are not balanced or whenever we try to use different, and often more logical, covariance structures.

First a bit of theory. Within **Proc Mixed** the **repeated** command plays a very important role in that it allows you to specify different covariance structures, which is something that you cannot do under **Proc GLM**. You should recall that in **Proc GLM** we assume that the covariance matrix meets our sphericity assumption and we go from there. In other words the calculations are carried out with the covariance matrix forced to sphericity. If that is not a valid assumption we are in trouble. Of course there are corrections due to Greenhouse and Geisser and Hyunh and Feldt, but they are not optimal solutions.

But what does compound symmetry, or sphericity, really represent? (The assumption is really about sphericity, but when speaking of mixed models most writers refer to compound symmetry, which is actually a bit more restrictive.) Most people know that compound symmetry means that the pattern of covariances or correlations is constant across trials. In other words, the correlation between trial 1 and trial 2 is equal to the correlation between trial 1 and trial 4 or trial 3 and trial 4, etc. But a more direct way to think about compound symmetry is to say that it requires that all subjects in each group change in the same way over trials. In other words the slopes of the lines regressing the dependent variable on time are the same for all subjects. Put that way it is easy to see that compound symmetry can really be an unrealistic assumption. If some of your subjects improve but others don't, you do not have compound symmetry and you make an error if you use a solution that assumes that you do. Fortunately **Proc Mixed** allows you to specify some other pattern for those covariances.

We can also get around the sphericity assumption using the MANOVA output from **Proc GLM**, but that too has its problems. Both standard univariate GLM and MANOVA GLM will insist on complete data. If a subject is missing even one piece of data, that subject is discarded. That is a problem because with a few missing observations we can lose a great deal of data and degrees of freedom.

**Proc Mixed** with repeated is different. Instead of using a least squares solution, which requires complete data, it uses a maximum likelihood solution, which does not make that assumption. (We will actually use a Restricted Maximum Likelihood (REML) solution.) When we have balanced data both least squares and REML will produce the same solution if we specify a covariance matrix with compound symmetry. But even with balanced data if we specify some other covariance matrix the solutions will differ. At first I am going to force sphericity by adding type = cs (which stands for compound symmetry) to the repeated statement. I will later relax that structure.

The first analysis below uses exactly the same data as for **Proc GLM**, though they are entered differently. Here data are entered in what is called "long form," as opposed to the "wide form" used for **Proc GLM**. This means that instead of having one line of data for each subject, we have one line of data for each observation. So with four measurement times we will have four lines of data for that subject.

Because we have a completely balanced design (equal sample sizes and no missing data) and because the time intervals are constant, the results of this analysis will come out exactly the same as those for **Proc GLM** so long as I specify type = cs. The data follow. I have used "card" input rather than reading a file just to give an alternative approach.

/*This will put output in C:\Documents and Settings\David Howell until I learn how to change it */ ods listing; ods html; data wicklong; input subj time group dv; timecont = time; cards;; /* The following lines plot the data */ Symbol1 I = join v = none r = 12; Proc gplot data = wicklong; Plot dv*time = subj/ nolegend; By group; Run; /* This is the main proc mixed procedure. */ proc mixed data = wicklong; class group subj time; model dv = group time group*time; repeated /subject = subj type = cs rcorr; run; ----------------------------------------------------------------------------------------

1 0 1.00 296.00 1 1 1.00 175.00 1 3 1.00 187.00 1 6 1.00 192.00 2 0 1.00 376.00 2 1 1.00 329.00 2 3 1.00 236.00 2 6 1.00 76.00 3 0 1.00 309.00 3 1 1.00 238.00 3 3 1.00 150.00 3 6 1.00 123.00 4 0 1.00 222.00 4 1 1.00 60.00 4 3 1.00 82.00 4 6 1.00 85.00 5 0 1.00 150.00 5 1 1.00 271.00 5 3 1.00 250.00 5 6 1.00 216.00 6 0 1.00 316.00 6 1 1.00 291.00 6 3 1.00 238.00 6 6 1.00 144.00 7 0 1.00 321.00 7 1 1.00 364.00 7 3 1.00 270.00 7 6 1.00 308.00 8 0 1.00 447.00 8 1 1.00 402.00 8 3 1.00 294.00 8 6 1.00 216.00 9 0 1.00 220.00 9 1 1.00 70.00 9 3 1.00 95.00 9 6 1.00 87.00 10 0 1.00 375.00 10 1 1.00 335.00 10 3 1.00 334.00 10 6 1.00 79.00 11 0 1.00 310.00 11 1 1.00 300.00 11 3 1.00 253.00 11 6 1.00 140.00 12 0 1.00 310.00 12 1 1.00 245.00 12 3 1.00 200.00 12 6 1.00 120.00 13 0 2.00 282.00 13 1 2.00 186.00 13 3 2.00 225.00 13 6 2.00 134.00 14 0 2.00 317.00 14 1 2.00 31.00 14 3 2.00 85.00 14 6 2.00 120.00 15 0 2.00 362.00 15 1 2.00 104.00 15 3 2.00 144.00 15 6 2.00 114.00 16 0 2.00 338.00 16 1 2.00 132.00 16 3 2.00 91.00 16 6 2.00 77.00 17 0 2.00 263.00 17 1 2.00 94.00 17 3 2.00 141.00 17 6 2.00 142.00 18 0 2.00 138.00 18 1 2.00 38.00 18 3 2.00 16.00 18 6 2.00 95.00 19 0 2.00 329.00 19 1 2.00 62.00 19 3 2.00 62.00 19 6 2.00 6.00 20 0 2.00 292.00 20 1 2.00 139.00 20 3 2.00 104.00 20 6 2.00 184.00 21 0 2.00 275.00 21 1 2.00 94.00 21 3 2.00 135.00 21 6 2.00 137.00 22 0 2.00 150.00 22 1 2.00 48.00 22 3 2.00 20.00 22 6 2.00 85.00 23 0 2.00 319.00 23 1 2.00 68.00 23 3 2.00 67.00 23 6 2.00 12.00 24 0 2.00 300.00 24 1 2.00 138.00 24 3 2.00 114.00 24 6 2.00 174.00

I have put the data in three columns to save space, but the real syntax statements would have 48 lines of data.

The first set of commands plots the results of each individual subject broken down by groups. Earlier we saw the group means over time. Now we can see how each of the subjects stands relative to the means of his or her group. In the ideal world the lines would start out at the same point on the Y axis (i.e. have a common intercept) and move in parallel (i.e. have a common slope). That isn't quite what happens here, but whether those are chance variations or systematic ones is something that we will look at later. We can see in the Control group that a few subjects decline linearly over time and a few other subjects, especially those with lower scores decline at first and then increase during follow-up.

Plots (Group 1 = Control, Group 2 = Treatment)

For **Proc Mixed** we need to specify that group, time, and subject are class variables. (See the syntax above.) This will cause SAS to treat them as factors (nominal or ordinal variables) instead of as continuous variables. The model statement tells the program that we want to treat group and time as a factorial design and generate the main effects and the interaction. (I have not appended a "/solution" to the end of the model statement because I don't want to talk about the parameter estimates of treatment effects at this point, but most people would put it there.) The repeated command tells SAS to treat this as a repeated measures design, that the subject variable is named "subj", and that we want to treat the covariance matrix as exhibiting compound symmetry, even though in the data that I created we don't appear to come close to meeting that assumption. The specification "rcorr" will ask for the estimated correlation matrix. (we could use "r" instead of "rcorr," but that would produce a covariance matrix, which is harder to interpret.)

The results of this analysis follow, and you can see that they very much resemble our analysis of variance approach using **Proc GLM**.

---------------------------------------------------------------------------------------- Proc Mixed with complete data. The Mixed Procedure Estimated R Correlation Matrix for subject 1 Row Col1 Col2 Col3 Col4 1 1.0000 0.4791 0.4791 0.4791 2 0.4791 1.0000 0.4791 0.4791 3 0.4791 0.4791 1.0000 0.4791 4 0.4791 0.4791 0.4791 1.0000 Covariance Parameter Estimates Cov Parm Subject Estimate CS subject 2539.36 Residual 2760.62 Fit Statistics -2 Res Log Likelihood 1000.8 AIC (smaller is better) 1004.8 AICC (smaller is better) 1004.9 BIC (smaller is better) 1007.2 Null Model Likelihood Ratio Test DF Chi-Square Pr > ChiSq 1 23.45 <.0001 Type 3 Tests of Fixed Effects Num Den Effect DF DF F Value Pr > F group 1 22 13.71 0.0012 time 3 66 45.14 <.0001 group*time 3 66 9.01 <.0001 ----------------------------------------------------------------------------------------

On this printout we see the estimated correlations between times. These are not the actual correlations, which appear below, but the estimates that come from an assumption of compound symmetry. That assumption says that the correlations have to be equal, and what we have here are basically average correlations. The actual correlations, averaged over the two groups using Fisher's transformation, are:

Estimated R Correlation Matrix for subj 1 Row Col1 Col2 Col3 Col4 1 1.0000 0.5695 0.5351 -0.01683 2 0.5695 1.0000 0.8612 0.4456 3 0.5351 0.8612 1.0000 0.4202 4 -0.01683 0.4456 0.4202 1.0000

Notice that they are quite different from the ones assuming compound symmetry, and that they don't look at all as if they fit that assumption. We will deal with this problem later. (I don't have a clue why the heading refers to "subject 1." It just does!)

There are also two covariance parameters. Remember that there are two sources of random effects in this design. There is our normal σ^{2}_{e}, which reflects random noise. In addition we are treating our subjects as a random sample, and there is thus random variance among subjects. Here I get to play a bit with expected mean squares. You may recall that the expected mean squares for the error term for the between-subject effect is E(MS_{w/in subj}) = σ_{e}^{2} + aσ_{π}^{2} and our estimate of σ_{e}^{2}, taken from the GLM analysis, is MS_{residual}, which is 2760.6218. The letter "a" stands for the number of measurement times = 4, and MS_{subj w/in grps} = 12918.0663, again from the GLM analysis. Therefore our estimate of σ_{π}^{2} = (12918.0663 + 2760.6218)/4 = 2539.36. These two estimates are our random part of the model and are given in the section headed Covariance Parameter Estimates. I don't see a situation in this example in which we would wish to make use of these values, but in other mixed designs they are useful.

You may notice one odd thing in the data. Instead of entering time as 1,2, 3, & 4, I entered it as 0, 1, 3, and 6. If this were a standard ANOVA it wouldn't make any difference, and in fact it doesn't make any difference here, but when we come to looking at intercepts and slopes, it will be very important how we designated the 0 point. We could have centered time by subtracting the mean time from each entry, which would mean that the intercept is at the mean time. I have chosen to make 0 represent the pretest, which seems a logical place to find the intercept. I will say more about this later.

I have just spent considerable time discussing a balanced design where all of the data are available. Now I want to delete some of the data and redo the analysis. This is one of the areas where mixed designs have an important advantage. I am going to delete scores pretty much at random, except that I want to show a pattern of different observations over time. It is easiest to see what I have done if we look at data in the wide form, so the earlier table is presented below with '.' representing missing observations. It is important to notice that data are missing completely at random, not on the basis of other observations.

Group Subj Time0 Time1 Time3 Time6 1 1 296 175 187 192 1 2 376 329 236 76 1 3 309 238 150 123 1 4 222 60 82 85 1 5 150 . 250 216 1 6 316 291 238 144 1 7 321 364 270 308 1 8 447 402 . 216 1 9 220 70 95 87 1 10 375 335 334 79 1 11 310 300 253 . 1 12 310 245 200 120 Group Subj Time0 Time1 Time3 Time6 2 13 282 186 225 134 2 14 317 31 85 120 2 15 362 104 . . 2 16 338 132 91 77 2 17 263 94 141 142 2 18 138 38 16 95 2 19 329 . . 6 2 20 292 139 104 . 2 21 275 94 135 137 2 22 150 48 20 85 2 23 319 68 67 . 2 24 300 138 114 174

If we treat this as a standard repeated measures analysis of variance, using **Proc GLM**, we have a problem. Of the 24 cases, only 17 of them have complete data. That means that our analysis will be based on only those 17 cases. Aside from a serious loss of power, there are other problems with this state of affairs. Suppose that I suspected that people who are less depressed are less likely to return for a follow-up session and thus have missing data. To build that into the example I could deliberately have deleted data from those who scored low on depression to begin with, though I kept their pretest scores. (I did not actually do this here.) Further suppose that people low in depression respond to treatment (or non-treatment) in different ways from those who are more depressed. By deleting whole cases I will have deleted low depression subjects and that will result in biased estimates of what we would have found if those original data points had not been missing. This is certainly not a desirable result.

To expand slightly on the previous paragraph, if we using **Proc GLM** , or a comparable procedure in other software, we have to assume that data are missing completely at random, normally abbreviated MCAR. (See Howell, 2008.) If the data are not missing completely at random, then the results would be biased. But if I can find a way to keep as much data as possible, and if people with low pretest scores are missing at one or more measurement times, the pretest score will essentially serve as a covariate to predict missingness. This means that I only have to assume that data are missing at random (MAR) rather than MCAR. That is a gain worth having. MCAR is quite rare in experimental research, but MAR is much more common. Using a mixed model approach requires only that data are MAR and allows me to retain considerable degrees of freedom. (That argument has been challenged by Overall & Tonidandel (2007), but in this particular example the data actually are essentially MCAR. I will come back to this issue later.)

The output from analyzing these data using **Proc GLM** follows. I give these results just for purposes of comparison, and I have omitted much of the printout.

----------------------------------------------------------------------------------------- Analysis of Wicksell missing data 25 13:39 Wednesday, March 31, 2011 The GLM Procedure Repeated Measures Analysis of Variance Tests of Hypotheses for Between Subjects Effects Source DF Type III SS Mean Square F Value Pr > F group 1 92917.9414 92917.9414 6.57 0.0216 Error 15 212237.4410 14149.1627 ---------------------------------------------------------------------------------------- The GLM Procedure Repeated Measures Analysis of Variance Univariate Tests of Hypotheses for Within Subject Effects Adj Pr > F Source DF Type III SS Mean Square F Value Pr > F G - G H - F time 3 238578.7081 79526.2360 32.42 <.0001 <.0001 <.0001 time*group 3 37996.4728 12665.4909 5.16 0.0037 0.0092 0.0048 Error(time) 45 110370.8507 2452.6856 Greenhouse-Geisser Epsilon 0.7386 Huynh-Feldt Epsilon 0.9300

Notice that we still have a group effect and a time effect, but the *F* for our interaction has been reduced by about half, and that is what we care most about. (In a previous version I made it drop to nonsignificant, but I relented here.) Also notice the big drop in degrees of freedom due to the fact that we now only have 17 subjects.

Now we move to the results using **Proc Mixed**. I need to modify the data file by putting it in its long form and to replacing missing observations with a period, but that means that I just altered 9 lines out of 96 (10% of the data) instead of 7 out of 24 (29%). The syntax would look exactly the same as it did earlier. The presence of "time' on the repeated statement is not necessary if I have included missing data by using a period, but it is needed if I just remove the observation completely. (At least that is the way I read the manual.) The results follow, again with much of the printout deleted:

Proc Mixed data = lib.wicklongMiss; class group time subject; model dv = group time group*time /solution; repeated time /subject = subject type = cs rcorr; run; ----------------------------------------------------------------------- Estimated R Correlation Matrix for subject 1 Row Col1 Col2 Col3 Col4 1 1.0000 0.4640 0.4640 0.4640 2 0.4640 1.0000 0.4640 0.4640 3 0.4640 0.4640 1.0000 0.4640 4 0.4640 0.4640 0.4640 1.0000 Covariance Parameter Estimates Cov Parm Subject Estimate CS subject 2558.27 Residual 2954.66 Fit Statistics -2 Res Log Likelihood 905.4 AIC (smaller is better) 909.4 AICC (smaller is better) 909.6 BIC (smaller is better) 911.8 Null Model Likelihood Ratio Test DF Chi-Square Pr > ChiSq 1 19.21 <.0001 ------------------------------------------------------------------------------------- Analysis with random intercept and random slope. Type 3 Tests of Fixed Effects Num Den Effect DF DF F Value Pr > F group 1 22 16.53 0.0005 time 3 57 32.45 <.0001 group*time 3 57 6.09 0.0011

This is a much nicer solution, not only because we have retained our significance levels, but because it is based on considerably more data and is not reliant on an assumption that the data are missing completely at random. Again you see a fixed pattern of correlations between trials which results from my specifying compound symmetry for the analysis.

To this point all of our analyses have been based on an assumption of compound symmetry. (The assumption is really about sphericity, but the two are close and **Proc Mixed** refers to the solution as type = cs.) But if you look at the correlation matrix given earlier it is quite clear that correlations further apart in time are distinctly lower than correlations close in time, which sounds like a reasonable result. Also if you looked at Mauchly's test of sphericity (not shown) it is significant with *p* = .012. While this is not a great test, it should give us pause. We really ought to do something about sphericity.

The first thing that we could do about sphericity is to specify that the model will make no assumptions whatsoever about the form of the covariance matrix. To do this I will ask for an unstructured matrix. This is accomplished by including type = un in the repeated statement. This will force SAS to estimate all of the variances and covariances and use them in its solution. The problem with this is that there are 10 things to be estimated and therefore we will lose degrees of freedom for our tests. But I will go ahead anyway. For this analysis I will continue to use the data set with missing data, though I could have used the complete data had I wished. I will include a request that SAS use procedures due to Hotelling-Lawley-McKeon (hlm) and Hotelling-Lawley-Pillai-Samson (hlps) which do a better job of estimating the degrees of freedom for our denominators. This is recommended for an unstructured model. The results are shown below.

Proc Mixed data = lib.WicksellLongMiss; class group time subject; model dv = group time group*time ; repeated time /subject = subject type = un hlm hlps rcorr; run; Estimated R Correlation Matrix for subject 1 Row Col1 Col2 Col3 Col4 1 1.0000 0.5858 0.5424 -0.02740 2 0.5858 1.0000 0.8581 0.3896 3 0.5424 0.8581 1.0000 0.3971 4 -0.02740 0.3896 0.3971 1.0000 Covariance Parameter Estimates Cov Parm Subject Estimate UN(1,1) subject 5548.42 UN(2,1) subject 3686.76 UN(2,2) subject 7139.94 UN(3,1) subject 2877.46 UN(3,2) subject 5163.81 UN(3,3) subject 5072.14 UN(4,1) subject -129.84 UN(4,2) subject 2094.43 UN(4,3) subject 1799.21 UN(4,4) subject 4048.07 Fit Statistics -2 Res Log Likelihood 883.7 AIC (smaller is better) 903.7 AICC (smaller is better) 906.9 BIC (smaller is better) 915.5 -------------------------------------------------------------------------------------------------- Same analysis but specifying an unstructured covariance matrix. The Mixed Procedure Null Model Likelihood Ratio Test DF Chi-Square Pr > ChiSq 9 40.92 <.0001 Type 3 Tests of Fixed Effects Num Den Effect DF DF F Value Pr > F group 1 22 17.95 0.0003 time 3 22 28.44 <.0001 group*time 3 22 6.80 0.0021 Type 3 Hotelling-Lawley-McKeon Statistics Num Den Effect DF DF F Value Pr > F time 3 20 25.85 <.0001 group*time 3 20 6.18 0.0038 ---------------------------------------------------------------------------------------- Same analysis but specifying an unstructured covariance matrix. The Mixed Procedure Type 3 Hotelling-Lawley-Pillai-Samson Statistics Num Den Effect DF DF F Value Pr > F time 3 20 25.85 <.0001 group*time 3 20 6.18 0.0038

Notice the matrix of correlations. From pretest to the 6 month follow-up the correlation with pretest scores has dropped from .46 to -.03, and this pattern is consistent. That certainly doesn't inspire confidence in compound symmetry.

The *F*s have not changed very much from the previous model, but the degrees of freedom for within-subject terms have dropped from 57 to 22, which is a huge drop. That results from the fact that the model had to make additional estimates of covariances. Finally, the hlm and hlps statistics further reduce the degrees of freedom to 20, but the effects are still significant. This would make me feel pretty good about the study if the data had been real data.

But we have gone from one extreme to another. We estimated two covariance parameters when we used type = cs and 10 covariance parameters when we used type = un. (Put another way, with the unstructured solution we threw up our hands and said to the program "You figure it out! We don't know what's going on.' There is a middle ground (in fact there are many). We probably do know at least something about what those correlations should look like. Often we would expect correlations to decrease as the trials in question are further removed from each other. They might not decrease as fast as our data suggest, but they should probably decrease. An autoregressive model, which we will see next, assumes that correlations between any two times depend on both the correlation at the previous time and an error component. To put that differently, your score at time 3 depends on your score at time 2 and error. (This is a first order autoregression model. A second order model would have a score depend on the two previous times plus error.) In effect an AR(1) model assumes that if the correlation between Time 1 and Time 2 is .51, then the correlation between Time 1 and Time 3 has an expected value of .512^{2} = .26 and between Time 1 and Time 4 has an expected value of .513^{3} = .13. Our data look reasonably close to that. (Remember that these are expected values of *r*, not the actual obtained correlations.) The solution using a first order autoregressive model follows.

Title 'Same analysis but specifying an autoregressive covariance matrix.'; Proc Mixed data = lib.WicksellLongMiss; class group subject time; model dv = group time group*time; repeated time /subject = subject type = AR(1) rcorr; run; Same analysis but specifying an autoregressive covariance matrix. Estimated R Correlation Matrix for subject 1 Row Col1 Col2 Col3 Col4 1 1.0000 0.6182 0.3822 0.2363 2 0.6182 1.0000 0.6182 0.3822 3 0.3822 0.6182 1.0000 0.6182 4 0.2363 0.3822 0.6182 1.0000 Covariance Parameter Estimates Cov Parm Subject Estimate AR(1) subject 0.6182 Residual 5350.25 Fit Statistics -2 Res Log Likelihood 895.1 AIC (smaller is better) 899.1 AICC (smaller is better) 899.2 BIC (smaller is better) 901.4 Null Model Likelihood Ratio Test DF Chi-Square Pr > ChiSq 1 29.55 <.0001 Type 3 Tests of Fixed Effects Num Den Effect DF DF F Value Pr > F group 1 22 17.32 0.0004 time 3 57 30.82 <.0001 group*time 3 57 7.72 <.0002

Notice the pattern of correlations. The .6182 as the correlation between adjacent trials is essentially an average of the correlations between adjacent trials in the unstructured case. The .3822 is just .61822^{2} and .2363 = .61823^{3}. Notice that tests on within-subject effects are back up to 57 df, which is certainly nice, and our results are still significant. This is a far nicer solution than we had using **Proc GLM**.

Now we have three solutions, but which should we choose? One aid in choosing is to look at the "Fit Statistics' that are printed out with each solution. These statistics take into account both how well the model fits the data and how many estimates it took to get there. Put loosely, we would probably be happier with a pretty good fit based on few parameter estimates than with a slightly better fit based on many parameter estimates. If you look at the three models we have fit for the unbalanced design you will see that the AIC criterion for the type = cs model was 909.4, which dropped to 903.7 when we relaxed the assumption of compound symmetry. A smaller AIC value is better, so we should prefer the second model. Then when we aimed for a middle ground, by specifying the pattern or correlations but not making SAS estimate 10 separate correlations, AIC dropped again to 899.1. That model fit better, and the fact that it did so by only estimating a variance and one correlation leads us to prefer that model.

You can accomplish the same thing using SPSS if you prefer. I will not discuss the syntax here, but the commands are given below. You can modify this syntax by replacing CS with UN or AR(1) if you wish. (A word of warning. For some reason SPSS has changed the way it reads missing data. In the past you could just put in a period and SPSS knew that was missing. It no longer does so. You need to put in something like -99 and tell it that -99 is the code for missing. While I'm at it, they changed something else. In the past it distinguished one value from another by looking for "white space." Thus if there were a tab, a space, 3 spaces, a space and a tab, or whatever, it knew that it had read one variable and was moving on to the next. NOT ANYMORE! I can't imagine why they did it, but for some ways of readig the data, if you put two spaces in your data file to keep numbers lined up vertically, it assumes that the you have skipped a variable. Very annoying. Just use one space or one tab between entries.)

MIXED dv BY Group Time /CRITERIA = CIN(95) MXITER(100) MXSTEP(5) SCORING(1) SINGULAR(0.000000000001) HCONVERGE(0, ABSOLUTE) LCONVERGE(0, ABSOLUTE) PCONVERGE(0.000001, ABSOLUTE) /FIXED = Group Time Group*Time | SSTYPE(3) /METHOD = REML /PRINT = DESCRIPTIVES SOLUTION /REPEATED = Time | SUBJECT(Subject) COVTYPE(CS) /EMMEANS = TABLES(Group) /EMMEANS = TABLES(Time) /EMMEANS = TABLES(Group*Time) .

The following commands will run the same analysis using the *R* program (or using S-PLUS). The results will not be exactly the same, but they are very close. Lines beginning with # are comments.

# Analysis of Wicksell Data with missing values data <- read.table(file.choose(), header = T) # WicksellLongMiss.dat attach(data) Time = factor(Time) Group = factor(Group) Subject = factor(Subject) library(lme4) model2 <- lmer(dv~Time + Group + Time:Group + (1|Subject)) summary(model2) anova(model2) model2@fixef model2@ranef # Now use the following model which allows for subjects to have different slopes over time. # Compare this with the SAS model with an unstructured correlation matrix. model3 <- lmer(dv~Time + Group + Time:Group + (Time|Subject)) summary(model3) anova(model3) # The following uses the nlme package, which was also written by Bates but # has been replaced by the newer mle4 package above. data <- read.table(file.choose(), header = T) attach(data) Time = factor(Time) Group = factor(Group) Subject = factor(Subject) library(nlme) model1 <- lme(dv ~ Time + Group + Time*Group, random = ~1 | Subject) summary(model1) anova(model1) # This model is very close to the one produced by SAS using compound symmetry, # when it comes to F values, and the log likelihood is the same. But the AIC # and BIC are quite different. The StDev for the Random Effects are the same # when squared. The coefficients are different because R uses the first level # as the base, whereas SAS uses the last.

In revising this version I found the following reference just stuck in the middle of nowhere. I don't recall why I did that, but Bodo Winter has an excellent page that I recommend that you look at. The link is http://www.bodowinter.com/tutorial/bw_LME_tutorial2.pdf.

This document is sufficiently long that I am going to create a new one to handle this next question. In that document we will look at other ways of doing much the same thing. The reason why I move to alternative models, even though they do the same thing, is that the logic of those models will make it easier for you to move to what are often called single-case designs or multiple baseline designs when we have finished with what is much like a traditional analysis of variance approach to what we often think of as traditional analysis of variance designs.

For the second part go to Mixed-Models-for-Repeated-Measures2.html

Guerin, L., and W.W. Stroup. 2000. A simulation study to evaluate PROC MIXED analysis of repeated measures data. p. 170-203. In Proc. 12th Kansas State Univ. Conf. on Applied Statistics in Agriculture. Kansas State Univ., Manhattan.

Howell, D.C. (2008) The analysis of variance. In Osborne, J. I., Best practices in Quantitative Methods. Sage.

Little, R. C., Milliken, G. A., Stroup, W. W., Wolfinger, R. D., & Schabenberger, O. (2006). SAS for Mixed Models. Cary. NC. SAS Institute Inc.

Maxwell, S. E. & Delaney, H. D. (2004) Designing Experiments and Analyzing Data: A Model Comparison Approach, 2nd edition. Belmont, CA. Wadsworth.

Overall, J. E., Ahn, C., Shivakumar, C., & Kalburgi, Y. (1999). Problematic formulations of SAS Proc.Mixed models for repeated measurements. Journal of Biopharmaceutical Statistics, 9, 189-216.

Overall, J. E. & Tonindandel, S. (2002) Measuring change in controlled longitudinal studies. British Journal of Mathematical and Statistical Psychology, 55, 109-124.

Overall, J. E. & Tonindandel, S. (2007) Analysis of data from a controlled repeated measurements design with baseline-dependent dropouts. Methodology, 3, 58-66.

Pinheiro, J. C. & Bates, D. M. (2000). Mixed-effects Models in S and S-Plus. Springer.

Some good references on the web are:

http://www.ats.ucla.edu/stat/sas/faq/anovmix1.htm

http://www.ats.ucla.edu/stat/sas/library/mixedglm.pdf

The following is a good reference for people with questions about using SAS in general.

http://ssc.utexas.edu/consulting/answers/sas/sas94.html

Downloadable Papers on Multilevel Models

Good coverage of alternative covariance structures

http://cda.morris.umn.edu/~anderson/math4601/gopher/SAS/longdata/structures.pdf

The main reference for SASProc Mixedis

Little, R.C., Milliken, G.A., Stroup, W.W., Wolfinger, R.D., & Schabenberger, O. (2006) SAS for mixed models, Cary, NC SAS Institute Inc.

See also

Maxwell, S. E. & Delaney, H. D. (2004). Designing Experiments and Analyzing Data (2nd edition). Lawrence Erlbaum Associates.

The classic reference for R is Penheiro, J. C. & Bates, D. M. (2000) Mixed-effects models in S and S-Plus. New York: Springer.

Last revised 6/28/2015