# Chapter 8 Repeated-measures ANOVA

In this Chapter, we will focus on performing repeated-measures ANOVA with R. We will use the same data analysed in Chapter 10 of SDAM, which is from an experiment investigating the “cheerleader effect.” The dataset is available in the sdamr package as cheerleader. We can load it from there, and inspect the first six cases, as usual:

library(sdamr)
head(cheerleader)
##   Participant Age    Sex                  Task LineClickAccuracy Excluded
## 1           1  47   Male Identical-Distractors         0.1361274        0
## 3           1  47   Male Identical-Distractors         0.1361274        0
## 4           1  47   Male Identical-Distractors         0.1361274        0
## 6           2  19 Female Identical-Distractors        -0.9752861        0
## 8           2  19 Female Identical-Distractors        -0.9752861        0
## 9           2  19 Female Identical-Distractors        -0.9752861        0
##   WhyExcluded                    Item Response
## 1        <NA>                   Alone 52.71289
## 3        <NA>           Control_Group 56.15966
## 4        <NA> Distractor_Manipulation 53.27871
## 6        <NA>                   Alone 52.47199
## 8        <NA>           Control_Group 55.29972
## 9        <NA> Distractor_Manipulation 55.17647

This dataset is a little messy, and includes participants who were excluded by the authors. So let’s first clean it up a little:

dat <- cheerleader
# remove participants which should be excluded
dat <- subset(dat, Excluded == 0)
# get rid of unused factor levels in Item by
dat$Item <- factor(dat$Item)

Another thing is that the labels of the factors don’t correspond to the ones I used in writing the SDAM chapter. Relabelling factors is somewhat tedious with base R. It’s easier to use the fct_recode function from the forcats package . This function takes a factor as its first argument, and then in the remaining arguments, you can specify a new label (unquoted) for existing labels (quoted). As usual, if you don’t have this package installed, you would first need to run install.packages("forcats") before running the code below:

dat$Presentation <- forcats::fct_recode(dat$Item, Different = "Control_Group", Similar = "Distractor_Manipulation")
dat$Version <- forcats::fct_recode(dat$Task, Identical = "Identical-Distractors", Variant = "Self-Distractors")

Let’s have a look at the resulting data.frame:

head(dat)
##   Participant Age    Sex                  Task LineClickAccuracy Excluded
## 1           1  47   Male Identical-Distractors         0.1361274        0
## 3           1  47   Male Identical-Distractors         0.1361274        0
## 4           1  47   Male Identical-Distractors         0.1361274        0
## 6           2  19 Female Identical-Distractors        -0.9752861        0
## 8           2  19 Female Identical-Distractors        -0.9752861        0
## 9           2  19 Female Identical-Distractors        -0.9752861        0
##   WhyExcluded                    Item Response Presentation   Version
## 1        <NA>                   Alone 52.71289        Alone Identical
## 3        <NA>           Control_Group 56.15966    Different Identical
## 4        <NA> Distractor_Manipulation 53.27871      Similar Identical
## 6        <NA>                   Alone 52.47199        Alone Identical
## 8        <NA>           Control_Group 55.29972    Different Identical
## 9        <NA> Distractor_Manipulation 55.17647      Similar Identical

Looks good! You can create a raincloud plot for this data as usual:

sdamr::plot_raincloud(data=dat, y=Response, groups = Presentation) + ggplot2::facet_wrap(~Version)

## 8.1 Long and wide data

The cheerleader data, and our dat data.frame is in the so-called long format. Data in the long format is structured so that each row contains is a single meaningful observation. Here, that translates to us having multiple rows for one participant (e.g. there are three rows for Participant 1). Data in the wide format has one row for each unit of observation (e.g. Participant). For some analyses, the long format is most suitable, whilst for others the wide format. It is therefore useful to be able to transform the data from one format to the other. This used to be a real pain back in the days when I started using R. Luckily, there are now tools available that make this a lot easier. Here, we will use the pivot_longer and pivot_wider functions from the tidyr package. The pivot_wider function is used to transform long-format data to the wide format. In the id_cols argument, you can list variable which identify a “unit of observation” (e.g. Participant), as well as other variables which don’t vary within subjects (such as condition). In the names_from argument, you can specify a variable which identifies the within-subjects levels, which is used to name the resulting new set of dependent variables. In the values_from argument, you specify the variable which contains the values of the new set of dependent variables:

wdat <- tidyr::pivot_wider(dat, id_cols = c("Participant", "Version"), names_from = Presentation, values_from = Response)
head(wdat)
## # A tibble: 6 × 5
##   Participant Version   Alone Different Similar
##   <fct>       <fct>     <dbl>     <dbl>   <dbl>
## 1 1           Identical  52.7      56.2    53.3
## 2 2           Identical  52.5      55.3    55.2
## 3 3           Identical  45.9      47.9    47.7
## 4 4           Identical  43.9      50.3    44.2
## 5 5           Identical  37.5      34.7    37.8
## 6 6           Identical  44.4      47.9    45.1

Note that there are quite a few variables no longer from this new wide-format data. This is not a problem, as we don’t need them for the present analyses (we could also have kept these in by including them in the id_cols argument). Also note that the class of this object is not a data.frame, but a tibble. A tibble is a “modern re-imagining of the data.frame” (https://tibble.tidyverse.org/). It is a central part of the tidyverse collection of R packages, which includes the tidyr and forcats packages we have just used, as well as the ggplot2 package, and many more. When you become more familiar with R programming, you will likely adopt more of the functions and principles of the tidyverse.

You can transform data from the wide format to the long format with the pivot_longer function. In the cols argument, you need to specify which columns in the wide format to transform into a single variable in the long format. In the names_to argument, you can specify the name of the resulting identifier for each value, and in the values_to argument, you can specify the name of the variable in the long format which contains the values:

ldat <- tidyr::pivot_longer(wdat, cols = c("Alone", "Different", "Similar"), names_to = "Presentation", values_to = "Response")
head(ldat)
## # A tibble: 6 × 4
##   Participant Version   Presentation Response
##   <fct>       <fct>     <chr>           <dbl>
## 1 1           Identical Alone            52.7
## 2 1           Identical Different        56.2
## 3 1           Identical Similar          53.3
## 4 2           Identical Alone            52.5
## 5 2           Identical Different        55.3
## 6 2           Identical Similar          55.2

## 8.2 Repeated-measures ANOVA with separate GLMs

In Chapter 10 of SDAM, we focused on performing repeated-measures ANOVA by constructing within-subjects composite scores, and then performing separate GLM analyses on these. We will start with this approach, and analyse the full 2 (Version: Identical, Variant) by 3 (Presentation: Alone, Different, Similar) design.

### 8.2.1 Computing within-subjects composite scores

The within-subjects composite scores are effectively contrasts, computed for each participant. Let’s define the following contrasts:

$$d_0$$ $$d_1$$ $$d_2$$
Alone 1 $$-\tfrac{2}{3}$$ $$0$$
Different 1 $$\tfrac{1}{3}$$ $$\tfrac{1}{2}$$
Similar 1 $$\tfrac{1}{3}$$ $$-\tfrac{1}{2}$$

We can compute each composite score from these contrasts as: $W_{j,i} = \frac{\sum_{k=1}^g d_{j,k} Y_{i,k}}{\sqrt{\sum_{k=1}^g d_{j,k}^2}}$ For $$W_0$$ (i.e. $$j=0$$), the computation in R is:

# compute the top part (numerator)
wdat$W0 <- wdat$Alone + wdat$Different + wdat$Similar
# apply scaling factor to get the correct SS
wdat$W0 <- wdat$W0/sqrt(3)

Similarly, we can compute $$W_1$$ and $$W_2$$ as:

wdat$W1 <- (1/3)*wdat$Different + (1/3)*wdat$Similar - (2/3)*wdat$Alone
wdat$W1 <- wdat$W1/sqrt((1/3)^2 + (1/3)^2 + (-2/3)^2)

wdat$W2 <- (1/2)*wdat$Different - (1/2)*wdat$Similar wdat$w2 <- wdat$W2/sqrt(2/4) ### 8.2.2 Performing a repeated-measures ANOVA with separate models We now have three new dependent variables ($$W_0$$, $$W_1$$, and $$W_2$$), and for each we can perform a GLM analysis. To do this, we need to set a suitable contrast for Version. As in SDAM, I will use $$(\tfrac{1}{2}, -\tfrac{1}{2})$$ for the Identical and Variant conditions respectively: contrasts(wdat$Version) <- c(0.5, -0.5)

We can then estimate a linear model for each composite variable. For $$W_0$$, we estimate:

mod0 <- lm(W0 ~ Version, data=wdat)
summary(mod0)
##
## Call:
## lm(formula = W0 ~ Version, data = wdat)
##
## Residuals:
##     Min      1Q  Median      3Q     Max
## -59.928  -4.593   1.922   9.059  31.259
##
## Coefficients:
##             Estimate Std. Error t value Pr(>|t|)
## (Intercept)  82.5764     2.1153  39.038   <2e-16 ***
## Version1      0.8647     4.2305   0.204    0.839
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
##
## Residual standard error: 16.23 on 57 degrees of freedom
## Multiple R-squared:  0.0007323,  Adjusted R-squared:  -0.0168
## F-statistic: 0.04177 on 1 and 57 DF,  p-value: 0.8388

Note that the estimated parameters are in the scale of $$W_0$$, not in the scale of the dependent variable ($$Y$$). We can get the rescaled estimates by dividing the estimates by the scaling factor ($$\sqrt{3}$$ in this case). The coefficients functions extracts the parameter estimates from the model. Hence, the rescaled estimates can be computed as:

coefficients(mod0)/sqrt(3)
## (Intercept)    Version1
##   47.675478    0.499217

To obtain equivalent $$F$$-tests, we can use the Anova function from the car package:

car::Anova(mod0, type=3)
## Anova Table (Type III tests)
##
## Response: W0
##             Sum Sq Df   F value Pr(>F)
## (Intercept) 401272  1 1523.9890 <2e-16 ***
## Version         11  1    0.0418 0.8388
## Residuals    15008 57
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

The procedure for the within-subjects composite scores $$W_1$$ and $$W_2$$ is similar. For $$W_1$$, the analysis is:

# Analysis for W1
mod1 <- lm(W1 ~ Version, data=wdat)
summary(mod1)
##
## Call:
## lm(formula = W1 ~ Version, data = wdat)
##
## Residuals:
##     Min      1Q  Median      3Q     Max
## -2.8388 -1.0661 -0.0307  0.9773  3.6227
##
## Coefficients:
##              Estimate Std. Error t value Pr(>|t|)
## (Intercept)  1.057739   0.190904   5.541 8.02e-07 ***
## Version1    -0.008033   0.381808  -0.021    0.983
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
##
## Residual standard error: 1.464 on 57 degrees of freedom
## Multiple R-squared:  7.766e-06,  Adjusted R-squared:  -0.01754
## F-statistic: 0.0004426 on 1 and 57 DF,  p-value: 0.9833
coefficients(mod1)/sqrt(6/9)
## (Intercept)    Version1
##  1.29546036 -0.00983829
car::Anova(mod1, type=3)
## Anova Table (Type III tests)
##
## Response: W1
##              Sum Sq Df F value    Pr(>F)
## (Intercept)  65.839  1 30.6992 8.023e-07 ***
## Version       0.001  1  0.0004    0.9833
## Residuals   122.245 57
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

And for $$W_2$$, it is:

# Analysis for W2
mod2 <- lm(W2 ~ Version, data=wdat)
summary(mod2)
##
## Call:
## lm(formula = W2 ~ Version, data = wdat)
##
## Residuals:
##      Min       1Q   Median       3Q      Max
## -2.02844 -0.47968 -0.06555  0.36866  2.52549
##
## Coefficients:
##             Estimate Std. Error t value Pr(>|t|)
## (Intercept)   0.1981     0.1184   1.673   0.0998 .
## Version1      0.5865     0.2368   2.476   0.0163 *
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
##
## Residual standard error: 0.9083 on 57 degrees of freedom
## Multiple R-squared:  0.09714,    Adjusted R-squared:  0.0813
## F-statistic: 6.133 on 1 and 57 DF,  p-value: 0.01626
coefficients(mod2)/sqrt(2/4)
## (Intercept)    Version1
##   0.2801500   0.8293672
car::Anova(mod2, type=3)
## Anova Table (Type III tests)
##
## Response: W2
##             Sum Sq Df F value  Pr(>F)
## (Intercept)  2.309  1  2.7991 0.09980 .
## Version      5.060  1  6.1330 0.01626 *
## Residuals   47.026 57
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1

Unfortunately, there is no simple way to obtain omnibus tests by combining these models. They can be computed “manually,” by extracting the relevant SSR, SSE, and df terms from the models. These can then be used to compute an $$F$$-statistic, and the $$p$$-value can then be computed by using the pf function.

This is not the most straightforward manner to obtain omnibus tests (the following sections show how to do this in a much more convenient manner). But to show it is doable, let’s compute the omnibus test for the main effect of Presentation in this way. The relevant SSR, SSE, and df terms are stored in the objects returned by the car::Anova function. We can see the structure of this object with the str function:

str(car::Anova(mod1, type=3))
## Classes 'anova' and 'data.frame':    3 obs. of  4 variables:
##  $Sum Sq : num 6.58e+01 9.49e-04 1.22e+02 ##$ Df     : num  1 1 57
##  $F value: num 3.07e+01 4.43e-04 NA ##$ Pr(>F) : num  8.02e-07 9.83e-01 NA
##  - attr(*, "heading")= chr [1:2] "Anova Table (Type III tests)\n" "Response: W1"

This shows that the car::Anova function returns a data.frame with the test results. The first row corresponds to the test of the intercept (which reflects the main effects of Presentation in this repeated-measures ANOVA). The last row contains the values for the error term. The structure for the mod2 analysis is the same. To get the relevant omnibus values, we can just take the appropriate elements from these data.frames. To get the omnibus SSR and $$\text{df}_1$$ terms, we can use:

SSR <- car::Anova(mod1, type=3)$"Sum Sq"[1] + car::Anova(mod2, type=3)$"Sum Sq"[1]
df1 <- car::Anova(mod1, type=3)$"Df"[1] + car::Anova(mod2, type=3)$"Df"[1]

And for the SSE and $$\text{df}_2$$ terms, we can use:

SSE <- car::Anova(mod1, type=3)$"Sum Sq"[3] + car::Anova(mod2, type=3)$"Sum Sq"[3]
df2 <- car::Anova(mod1, type=3)$"Df"[3] + car::Anova(mod2, type=3)$"Df"[3]

With these values, the $$F$$-statistic can then be computed as follows:

Fstat <- (SSR/df1)/(SSE/df2)

Finally, the $$p$$-value can be obtained as

1-pf(Fstat, df1=df1, df2=df2)
## [1] 4.214718e-09

Note that we need to use 1-pf as the pf function computes the probability $$P(F \leq \text{value})$$, whilst we need $$P(F > \text{value})$$, and this equals $$P(F > \text{value}) = 1 - P(F \leq \text{value})$$.

The steps we have just taken is a perfectly valid manner to conduct a repeated-measures ANOVA, but it is a laborious process. An easier way to conduct repeated-measures ANOVA is provided in the car or afex package. Neither of these packages provide the tests for the individual contrasts we have just obtained. But these can be computed with the emmeans package, after conducting the omnibus tests.

## 8.3 Repeated-measures ANOVA with the car package

When you have data in the wide format, you can obtain a repeated-measures ANOVA by using the Anova function from the car package . As you will see later, the analysis is more straightforward with the afex package, but this requires data to be in the long format.

The first step to performing a repeated-measures ANOVA with the car package is to perform a linear model for a multivariate dependent variable, which basically means providing a matrix of each repeated measurement as the DV. This is done by collating the variables within a cbind (for column-bind) argument within the model formula:

mvmod <- lm(cbind(Alone, Different, Similar) ~ Version, data=wdat)

So here, we are modelling the Alone, Different, and Similar attractiveness ratings simultaneously as a function of the Version categorical predictor. This model is basically a set of three linear regressions, as you can see from the output:

mvmod
##
## Call:
## lm(formula = cbind(Alone, Different, Similar) ~ Version, data = wdat)
##
## Coefficients:
##              Alone     Different  Similar
## (Intercept)  46.81184  48.30539   47.90920
## Version1      0.50578   1.08239   -0.09051

The next step is to construct an object which reflects the structure of these three measurements. This has to be done with a separate data.frame, with one row for each variable included in the cbind function specifying the multivariate DV. In this case, there is a single categorical predictor underlying all three measurements. So our data.frame can contain a single factor:

idata <- data.frame(Presentation = factor(c("Alone", "Different", "Similar")))
idata
##   Presentation
## 1        Alone
## 2    Different
## 3      Similar

The next step is to supply a useful contrast for this within-subjects factor:

contrasts(idata$Presentation) # check the levels ## Different Similar ## Alone 0 0 ## Different 1 0 ## Similar 0 1 contrasts(idata$Presentation) <- cbind(c(-2/3, 1/3, 1/3), c(0,1/2, -1/2))
contrasts(idata$Presentation) ## [,1] [,2] ## Alone -0.6666667 0.0 ## Different 0.3333333 0.5 ## Similar 0.3333333 -0.5 With these elements in place, we are finally ready to perform the repeated-measures ANOVA. This involves calling the car::Anova function with the multivariate linear model as the first argument, and supplying the within-subjects structure through the idata argument. Additionally, you need to supply a right-hand-sided formula in the idesign argument in order to specify which effects to include as within-subjects factors. The type=3 argument, as usual, specifies we would like to perform Type-3 SS tests. rmaov <- car::Anova(mvmod, idata=idata, idesign = ~Presentation, type=3) rmaov ## ## Type III Repeated Measures MANOVA Tests: Pillai test statistic ## Df test stat approx F num Df den Df Pr(>F) ## (Intercept) 1 0.96395 1523.99 1 57 < 2.2e-16 *** ## Version 1 0.00073 0.04 1 57 0.83878 ## Presentation 1 0.36601 16.17 2 56 2.872e-06 *** ## Version:Presentation 1 0.09730 3.02 2 56 0.05691 . ## --- ## Signif. codes: 0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1 By default, because we have used a multivariate DV, this will show a so-called MANOVA (Multivariate ANalysis of VAriance). To obtain an ANOVA, we need to set the multivariate argument in the summary function to FALSE: summary(rmaov, multivariate=FALSE) ## Warning in summary.Anova.mlm(rmaov, multivariate = FALSE): HF eps > 1 treated as ## 1 ## ## Univariate Type III Repeated-Measures ANOVA Assuming Sphericity ## ## Sum Sq num Df Error SS den Df F value Pr(>F) ## (Intercept) 401272 1 15008.3 57 1523.9890 < 2.2e-16 *** ## Version 11 1 15008.3 57 0.0418 0.83878 ## Presentation 70 2 216.3 114 18.5675 1.047e-07 *** ## Version:Presentation 10 2 216.3 114 2.6670 0.07379 . ## --- ## Signif. codes: 0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1 ## ## ## Mauchly Tests for Sphericity ## ## Test statistic p-value ## Presentation 0.98188 0.59925 ## Version:Presentation 0.98188 0.59925 ## ## ## Greenhouse-Geisser and Huynh-Feldt Corrections ## for Departure from Sphericity ## ## GG eps Pr(>F[GG]) ## Presentation 0.9822 1.314e-07 *** ## Version:Presentation 0.9822 0.07485 . ## --- ## Signif. codes: 0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1 ## ## HF eps Pr(>F[HF]) ## Presentation 1.016935 1.046537e-07 ## Version:Presentation 1.016935 7.379140e-02 In addition to an ANOVA table which contains the omnibus tests for the within- and between-subjects effects, the output provides the Mauchly sphericity test, and subsequently the Greenhouse-Geisser and Huynh-Feldt corrected tests. The tables corresponding to these latter two corrected tests report the Greenhouse-Geisser and Huynh-Feldt estimates (as GG eps and HF eps respectively) of what I have denoted as $$\hat{\zeta}$$, but is more commonly denoted as $$\hat{\epsilon}$$, and the $$p$$-value which results from applying the correction to the degrees of freedom reported in the Univariate Type III Repeated-Measures ANOVA Assuming Sphericity table. A notable absence is the tests of the specific contrasts. We can obtain these by performing analyses on the within-subjects composite scores, as we did in the previous section. ## 8.4 Repeated-measures ANOVA with the afex package The afex package provides a convenient interface to the car::Anova() function, via its afex::aov_car() function. To use this function, the data needs to be in the long format. You can specify the model with the usual formula interface, and you don’t need to worry about a multivariate response and such things. There is one new thing, however: To specify a repeated-measures ANOVA, the formula needs to contain a special Error() argument. Within the Error argument, you first state the variable which identifies the “units of observations” (i.e. Participant in this case). Then, after a forward-slash (“/”), you list the repeated-measures factor(s). So, the way to perform the repeated-measures ANOVA with the afex package, and the long data (ldat) we created earlier, is: afmod <- afex::aov_car(Response ~ Version*Presentation + Error(Participant/Presentation), data=ldat) ## Contrasts set to contr.sum for the following variables: Version afmod ## Anova Table (Type 3 tests) ## ## Response: Response ## Effect df MSE F ges p.value ## 1 Version 1, 57 263.30 0.04 <.001 .839 ## 2 Presentation 1.96, 111.97 1.93 18.57 *** .005 <.001 ## 3 Version:Presentation 1.96, 111.97 1.93 2.67 + <.001 .075 ## --- ## Signif. codes: 0 '***' 0.001 '**' 0.01 '*' 0.05 '+' 0.1 ' ' 1 ## ## Sphericity correction method: GG afex provides an abridged ANOVA table, where the Greenhouse-Geisser correction is automatically applied. Note that afex automatically sets contrasts to contr.sum. That is useful here, as we haven’t set the contrast for Version in the ldat data. Because neither afex, nor the car::Anova package on which it relies, provides parameter estimates for the GLM, it doesn’t really matter whether you supply your own (sum-to-zero) contrasts, or whether you let afex pick a contr.sum() contrast for you. The afex package provides a convenient wrapper around the car::Anova() function, and saves you a lot of work if you have data in the long format. I would generally recommend storing data in the long format, as this also makes it easier to apply linear mixed-effects models. You can obtain the results as reported by the car::Anova() function by calling the summary function: summary(afmod) ## Warning in summary.Anova.mlm(object$Anova, multivariate = FALSE): HF eps > 1
## treated as 1
##
## Univariate Type III Repeated-Measures ANOVA Assuming Sphericity
##
##                      Sum Sq num Df Error SS den Df  F value    Pr(>F)
## (Intercept)          192932      1  15008.3     57 732.7366 < 2.2e-16 ***
## Version                  11      1  15008.3     57   0.0418   0.83878
## Presentation             45      2    216.3    114  11.7553 2.283e-05 ***
## Version:Presentation     10      2    216.3    114   2.6670   0.07379 .
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
##
##
## Mauchly Tests for Sphericity
##
##                      Test statistic p-value
## Presentation                0.98188 0.59925
## Version:Presentation        0.98188 0.59925
##
##
## Greenhouse-Geisser and Huynh-Feldt Corrections
##  for Departure from Sphericity
##
##                      GG eps Pr(>F[GG])
## Presentation         0.9822   2.62e-05 ***
## Version:Presentation 0.9822    0.07485 .
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '.' 0.1 ' ' 1
##
##                        HF eps   Pr(>F[HF])
## Presentation         1.016935 2.282549e-05
## Version:Presentation 1.016935 7.379140e-02

The afex package also has a function, called nice, to display the abbreviated ANOVA table we saw earlier. The arguments of this function allow you to change various aspects of the displayed results. For instance, by default, the “generalized eta-square” is used as a measure of effect size. You can change this to the partial eta-square by setting es="pes". You can also change the correction to the degrees of freedom, from the default correction = "GG" (Greenhouse-Geisser), by setting correction = "HF" (Huynh-Feldt) or correction = "none" (no correction). So, for example, we might use

afex::nice(afmod, es="pes", correction = "none")
## Anova Table (Type 3 tests)
##
## Response: Response
##                 Effect     df    MSE         F   pes p.value
## 1              Version  1, 57 263.30      0.04 <.001    .839
## 2         Presentation 2, 114   1.90 11.76 ***  .171   <.001
## 3 Version:Presentation 2, 114   1.90    2.67 +  .045    .074
## ---
## Signif. codes:  0 '***' 0.001 '**' 0.01 '*' 0.05 '+' 0.1 ' ' 1

## 8.5 Contrasts with the emmeans package

To obtain the individual contrast estimates and tests for a repeated-measures ANOVA, perhaps the most straightforward procedure is via the emmeans package . We have already discussed the use of this package in Section 7.4. The emmeans::emmeans() function calculates estimated marginal means, and it can do so for objects that are returned by the afex::aov_car() function. For example, we can obtain the marginal means for the different levels of Version as:

em_version <- emmeans::emmeans(afmod, specs = ~ Version)

The specs argument should contain a right-sided formula with the factor(s) for which you want to compute the marginal means. You can see that the emmeans function computes the estimated marginal means, their standard error, associated degrees of freedom, and confidence intervals:

em_version
##  Version   emmean   SE df lower.CL upper.CL
##  Identical   47.9 1.77 57     44.4     51.5
##  Variant     47.4 1.68 57     44.1     50.8
##
## Results are averaged over the levels of: Presentation
## Confidence level used: 0.95

Contrasts are really just differences between (sets of) marginal means, or even more generally, linear functions of marginal means. By applying a contrast code to the marginal means, we obtain a new value (e.g. a difference between marginal means), which also comes with a standard error and associated degrees of freedom. These can be used in a one-sample $$t$$-test, to test whether this new values is equal to an assumed value (e.g. 0). For example, we may want to test whether the difference between the marginal mean of the Identical condition and the Variant condition is equal to 0, i.e. $$H_0: \mu_I - \mu_V = 0$$. The contrast to transform the means into this difference is $$(1,-1)$$, because the sum of the means multiplied by these values is $$1 \times \mu_I + (-1) \times \mu_V = \mu_I - \mu_V$$. Note that these contrasts work slightly differently than when using contrast-coded predictors in the GLM. There, we would have used the values $$(\tfrac{1}{2}, -\tfrac{1}{2})$$ to obtain the same contrast. That is because the slopes of orthogonal contrast-coded predictors are $\beta_j = \frac{\sum_{k=1}^g c_{j,k} \times \mu_k}{\sum_{k=1}^g c^2_{j,k}}$ and we would aim for this slope to represent $$\mu_I - \mu_V$$. Using $$(\tfrac{1}{2}, -\tfrac{1}{2})$$, the slope would be exactly this: $\frac{ \tfrac{1}{2} \mu_I - \tfrac{1}{2} \mu_V}{(\tfrac{1}{2})^2 + (-\tfrac{1}{2})^2} = \frac{ \tfrac{1}{2} \mu_I - \tfrac{1}{2} \mu_V}{\tfrac{1}{2}} = \mu_I - \mu_V$ As the contrasts in the emmeans package apply directly to the means, we don’t have to worry about the $$\sum_{k=1}^g c^2_{j,k}$$ term which scales the slopes of contrast-coded predictors in the GLM. That is why, when using the emmeans package, we can use $$(1,-1)$$ as our contrast, instead of $$(\tfrac{1}{2}, -\tfrac{1}{2})$$.

The emmeans::contrast() function allows us to compute such contrasts of marginal means, and the corresponding one-sample $$t$$-test (against the null-hypothesis that the resulting value is equal to 0). In the methmod argument, we here supply a named list. The name is not necessary, but is helpful in identifying the contrasts when you test multiple. The key thing is that we supplied a contrast with c(1,-1). The output provides the estimate of the difference between $$\mu_I$$ and $$\mu_V$$, as well as a $$t$$-test and $$p$$-value:

emmeans::contrast(em_version, method=list("I - V" = c(1,-1)))
##  contrast estimate   SE df t.ratio p.value
##  I - V       0.499 2.44 57   0.204  0.8388
##
## Results are averaged over the levels of: Presentation

We can see that this test is not significant, and hence we can’t reject the null-hypothesis that the true difference between the means is equal to 0. The results are identical to the test of the slope of Version1 for the model of W0 in Section 8.2.

We can follow the same procedure to tests contrasts for the Presentation factor. In this case, the list of contrasts specified in the method argument has two elements. In the first contrast, we want to determine the difference $\frac{\mu_D + \mu_S}{2} - \mu_A$ which we can do through the contrast c(-1,.5,.5) (applied to the A, D, and S conditions respectively). In the second contrast, we want to determine the difference $\mu_D - \mu_S$ which we can do through the contrast c(0,1,-1). The following code computes the marginal means and then performs the contrast-tests on these:

em_presentation <- emmeans::emmeans(afmod, specs = ~ Presentation)
emmeans::contrast(em_presentation, method=list("(D + S)/2 - A" = c(-1,.5,.5), "D - S" = c(0,1,-1)))
##  contrast      estimate    SE df t.ratio p.value
##  (D + S)/2 - A    1.295 0.234 57   5.541  <.0001
##  D - S            0.396 0.237 57   1.673  0.0998
##
## Results are averaged over the levels of: Version

This replicates the earlier results we obtained in Section 8.2 from the tests of the intercepts of W1 and W2.

Finally, we can also consider the marginal means of the combinations of the Presentation and Version factors. We do this by specifying the full (main effects and interaction) model in the specs argument:

em_pv <- emmeans::emmeans(afmod, specs = ~ Presentation*Version)

The marginal means are

em_pv
##  Presentation Version   emmean   SE df lower.CL upper.CL
##  Alone        Identical   47.1 1.78 57     43.5     50.6
##  Different    Identical   48.8 1.78 57     45.3     52.4
##  Similar      Identical   47.9 1.80 57     44.3     51.5
##  Alone        Variant     46.6 1.69 57     43.2     49.9
##  Different    Variant     47.8 1.69 57     44.4     51.1
##  Similar      Variant     48.0 1.71 57     44.5     51.4
##
## Confidence level used: 0.95

An interaction implies that a contrast for one experimental factor is moderated by the levels of another experimental factor. For example, the difference between $\frac{\mu_D + \mu_S}{2} - \mu_A$ might be different in the Identical compared to the Variant condition. If that is the case, then the difference between these differences would not equal 0. Such a “difference of differences” is most clearly stated in an equation: $\left(\frac{\mu_{I,D} + \mu_{I,S}}{2} - \mu_{I,A}\right) - \left(\frac{\mu_{V,D} + \mu_{V,S}}{2} - \mu_{V,A}\right)$ If we were to write this as a sum of all six means, we would do so as follows: $\tfrac{1}{2} \times \mu_{I,D} + \tfrac{1}{2} \mu_{I,S} + (-1) \times \mu_{I,A} + (-\tfrac{1}{2}) \times \mu_{V,D} + (-\tfrac{1}{2}) \times \mu_{V,S} + 1 \times \mu_{V,A}$ Hence, the implied contrast code is c(.5, .5, -1, -.5, -.5, 1). Following a similar logic, the implied contrast code for the second interaction is c(0, 1, -1, 0, -1, 1). Supplying these contrast codes to the emmeans::contrast function, we obtain the following results:

emmeans::contrast(em_pv, method=list("c1 by d1" = c(-1, .5, .5, 1, -.5, -.5), "c1 by d2" = c(0, 1, -1, 0, -1, 1)))
##  contrast estimate    SE df t.ratio p.value
##  c1 by d1 -0.00984 0.468 57  -0.021  0.9833
##  c1 by d2  1.17290 0.474 57   2.476  0.0163

This replicates the results of the estimates and tests of the slopes of Version1 in the models of W1 and W2 of Section 8.2 exactly.

So it doesn’t really matter all that much what form of contrast coding you use in the original analysis (as long as you use a form of sum-to-zero contrast coding). You can always perform the contrast tests afterwards by using the emmeans package (and other packages which provide similar functionality).

### References

Fox, John, Sanford Weisberg, and Brad Price. 2021. Car: Companion to Applied Regression. https://CRAN.R-project.org/package=car.
Lenth, Russell V. 2021. Emmeans: Estimated Marginal Means, Aka Least-Squares Means. https://github.com/rvlenth/emmeans.
Singmann, Henrik, Ben Bolker, Jake Westfall, Frederik Aust, and Mattan S. Ben-Shachar. 2021. Afex: Analysis of Factorial Experiments. https://CRAN.R-project.org/package=afex.
Wickham, Hadley. 2021a. Forcats: Tools for Working with Categorical Variables (Factors). https://CRAN.R-project.org/package=forcats.
———. 2021b. Tidyr: Tidy Messy Data. https://CRAN.R-project.org/package=tidyr.