partR2: partitioning R2 in generalized linear mixed models

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Bioinformatics and Genomics


Coefficients of determination R2 are of interest in the study of ecology and evolution, because they quantify the amount of variation explained by a linear model (Edwards et al., 2008). By doing so, they go beyond significance testing in putting effects in perspective of the phenotypic variance. R2 is expressed as a proportion of the total variance in the response, which represents a biologically relevant quantity if the total variation is representative for the total population (De Villemereuil et al., 2018). The total coefficient of determination in a generalised linear mixed model (GLMM) quantifies the variance explained by all fixed effects together (marginal R2 sensu Nakagawa & Schielzeth, 2013, also known as the total correlation coefficient, Watanabe, 1960).

However, it is often of interest to attribute explained variation to individual predictors. Semi-partial coefficients of determination, also known as part R2, decompose the variance of R2 into components uniquely explained by individual predictors (Jaeger et al., 2017; Jaeger, Edwards & Gurka, 2019) or sets of predictors (Fig. 1). The set of all predictors in the model yields the total proportion of variance explained by the fixed part of the model (total R2). With correlations among predictors, it often happens that predictors in univariate regressions explain a large share of the variance, but do not show large part R2 if other correlated predictors are included in the model. Note that part R2 estimates the proportion of the variance in the response explained by a predictor while accounting for covariance between this predictor and the other predictors in the model, whereas the (arguably more familiar) partial R2 estimates the proportion of the variance that is explained by a predictor of interest after accounting for the other predictors from the response as well as the predictor of interest. The difference is subtle, but important (see more below). Therefore, part R2 represents ‘variance accounted for’ in relation to the total variance, but partial R2 does not. Consequently, part R2 will be conceptually easier to compare with (total) R2.

Conceptual framework for the estimation of proportions of variance components in a mixed model.

Figure 1: Conceptual framework for the estimation of proportions of variance components in a mixed model.

The large grey circle symbolizes the variance in a response Y, the dark grey area on the top indicates the share explained by random effects and the coloured ellipses symbolize variance in covariates with intersections indicating jointly explained variances. partR2 calculates total R2, part R2 for individual predictors and sets of predictors as well as inclusive R2. The package does not report partial R2 and commonalities, although they could be calculated from the partR2 output.

Structure coefficients provide a valuable addition to part R2 in the decomposition of the phenotypic variance (Nimon et al., 2008; Yeatts et al., 2017). Structure coefficients quantify the correlation between individual predictors and the linear predictor. Predictors that correlate well with a response, but are fitted with collinear predictors may show large structure coefficients as they are correlated to the predicted response, but low part R2 as other predictors explain part of the same variance. Structure coefficients range from −1 to 1 with their absolute value expressing the correlation relative to a perfect correlation if a single predictor explains as much as the total fixed part of the model.

Structure coefficients are correlations and since the square of a correlation yields the variance explained, we can use structure coefficients to estimate the total variance explained by a predictor (Nimon et al., 2008). We call this the inclusive R2 of a predictor and calculate it as the squared structure coefficient, i.e., its contribution to the linear predictor independent of other predictors (Nimon et al., 2008) times the proportion of variance explained by the linear predictor (which is the ‘total’ marginal R2 of the model) (see also Nathans, Oswald & Nimon, 2012). As far as we are aware, inclusive R2 has not been implemented before, but it provides valuable insights into the structure of the variance explained (Fig. 1).

Here, we introduce partR2, a versatile package for estimating part R2, inclusive R2, structure coefficients and beta weights from mixed-effects models. Figure 1 gives an overview of how variances are calculated and how they relate to partial R2 and to commonality analysis (Ray Mukherjee et al., 2014; Seibold & McPhee, 1979; Zientek & Thompson, 2006). We illustrate how to use partR2 with real example datasets for Gaussian and binomial GLMMs, discuss how to estimate part R2in the presence of interactions and discuss some challenges and limitations. The Landesamt fur Natur, Umwelt und Verbraucherschutz Nordrhein-Westfalen ”LANUV NRW” (Germany) approved this research (reference number: 84-02.04.2015.A439).

Mathematical representation

Part R2

A Gaussian mixed-effects model can be written as: y = X β + α k + ε α k N 0 , σ α k 2 ε N 0 , σ ε 2

Where y is a vector of response values (outcomes), X is the design matrix of fixed effects, β is a vector of regression coefficients, ∑αk is the random part of the model that might contain multiple random effects and ε is a vector of residual deviations. The linear predictor η represents the vector of predicted values from the fixed part of the model as η = Xβ. Note that we dealing with estimates of regression coefficients and variance components throughout (hence all β should be read as β ̂ ).

Since we are interested in the proportion of the phenotypic variance explained, we symbolize variance components by upper case Y and index by the source of variance (as in Fig. 1). While variances are frequently represented as V with the source of variance as an index, this leads to ambiguity for VX which might represent variance in y explained by x or variance in x itself, which is why we use this alternative notation. The total variance in the response is YTotal = var(y) and is estimated from the raw data or from the model (see below). The variance of the residuals is estimated by the model as YR = var(ε). The variance of the (sum of) random effects is estimated by the model as YRE = var(∑αk) and the variance explained by fixed effects can be estimated as the variance in the linear predictor YX = var().

The coefficient of determination R2 estimates the proportion of variance in the response that is explained by fixed effects. The coefficient of determination R2 for the total fixed part of the model is thus: R X 2 = Y X Y X + Y R E + Y R = Y X Y T o t a l

Note that the sum of the components in the denominator might deviate numerically from the total outcome variance in the raw data. However, conceptually they are the same in that they represent the population-level outcome variance. The variance in the outcome is an estimate from the specific sample, while the sum of components of the mixed model represents a population-level estimate given the data and the model.

A reduced model with a (set of) fixed effect predictors X removed but the same random effect structure can be fitted as (now using the tilde to highlight the differences from Eq. (1)): y = X ˜ β ˜ + α ˜ k + ε ˜ α ˜ k N 0 , σ α ˜ k 2 ε ˜ N 0 , σ ε ˜ 2 with the variance in the linear predictor of the reduced model being Y X ˜ = v a r X β .

The variance uniquely explained by X is then the difference between the variance explained by fixed effects in the full and the reduced model Y X = Y X Y X ˜ . Part R2 sets this variance in proportion to the total outcome variance: R X 2 = Y X Y X ˜ Y X + Y R E + Y R = Y X Y X ˜ Y T o t a l

The process of fitting a reduced model, estimation of YX and estimation of R X 2 can be repeated for all predictors and combinations of predictors. At the limit for a model with all fixed effects removed, R X 2 = R X 2 .

Side-note on partial R2

For completeness we note that the partial R2 could be calculated as: R X 2 = Y X Y X ˜ Y T o t a l Y X ˜

However, this estimate does not put the explained variance in perspective of the total variance in the response. It has the major disadvantage that the denominator depends on Y X ˜ . The same effect in terms of YX thus appears larger if the reduced model explains more variance (larger Y X ˜ ). Even in the case of independent additive predictors, the contributions of the different fixed effects do not sum up to R X 2 , because of the change in the denominator that different YX are compared to. Finally, since we are interested in explaining phenotypic variation in some biological response (the phenomenon to be explained), we think that part R2 is the more relevant quantity, as it represents the proportion of variance in the response uniquely explained by X.

Inclusive R2

Structure coefficients are the Pearson correlations between a particular predictor of interest x and the linear predictor η. Note that we now use a lower case x to indicate that we are dealing with a single predictor. Structure coefficients are quantified from the full model as: S C x = cor η , x

The squared correlation between two variables a and b gives the variance explained for these variables cor a , b 2 = R a 2 . The squared structure correlations thus quantify the proportion of variance in the linear predictor YX that is explained by a the predictor of interest x. Since the proportion of outcome variance explained by the linear predictor in the full model is R X 2 , the inclusive variance explained by predictor x is: I R x 2 = S C 2 R X 2

Inclusive R2 as we define it here, complements part R2 by giving additional insights. While part R2 quantifies the variance uniquely explained by a predictor (or set of predictors), inclusive R2 quantifies the total proportion of variance explained in the model, both uniquely and jointly with other predictors. In the special case of a single predictor in a model SCx = cor(ηx) = 1, such that I R x 2 = R X 2 .

Part R2 in non-Gaussian models

For Gaussian models there is a single residual error term ε with variance YR = var(ε). For non-Gaussian models, however, there is additional error that arises from the link function that translates latent-level predictions to observed outcomes. This variance can be approximated for a variety of link functions and error distributions (Nakagawa & Schielzeth, 2010; Nakagawa, Johnson & Schielzeth, 2017). Our R package currently implements distribution-specific variances for Poisson models with log and square root link functions and binomial models with logit and probit link functions. For Poisson models and non-binary binomial models (proportion models), partR2 also fits an observational level random effect (if none is fitted already) to estimate variance due to overdispersion (Harrison, 2014). Both the overdispersion variance, now denoted YR and the distribution-specific variance YD are included in the denominator of the part R2 calculation: R X 2 = Y X Y X ˜ Y X + Y R E + Y D + Y R

Notably, there are other estimation methods for R2 for non-Gaussian models or GLMM (Jaeger et al., 2017; Piepho, 2019). Currently, partR2 only implements the method based on Nakagawa & Schielzeth (2013) and Nakagawa, Johnson & Schielzeth (2017).

Other implementations in R packages

There are a few R packages that calculate part R2 for linear models (lm), for example rockchalk::getDeltaRsquare (Johnson & Grothendieck, 2019). Other packages calculate partial R2 (not part R2) such as asbio::partial.R2 (Aho, 2020) and rr2::R2 (Ives & Li, 2018) for linear models and rsq::rsq.partial (Zhang, 2020) for linear models and generalized linear models (glm). Note that partial R2 is different from part (semi-partial) R2 (partial R2 >part R2), since it represents the unique variance explained by a particular predictor but after removing (‘partialling out’) the variance explained by the other predictors (Yeatts et al., 2017, Fig. 1). The ppcor package calculates semi-partial and partial correlations, but does not work on fitted GLM or GLMM models (Kim, 2015). The package yhat features functions for commonality analyses in glms (Nimon, Oswald & Roberts, 2020). None of these packages estimates part R2 for mixed-effects models that we focus on here.

Several packages estimate (marginal) R2 as the variance explained by all fixed effects in linear mixed-effects models. This includes performance::r2_nakagawa (Lüdecke et al., 2020), MuMIn::r.squaredGLMM (Bartoń, 2019), and rptR::rpt (Stoffel, Nakagawa & Schielzeth, 2017). These packages do not allow to estimate part R2. The only versatile package to estimate part R2 from linear mixed-models is r2glmm (Jaeger, 2017). The function r2glmm::r2beta computes part R2 from lmer, lme and glmmPQL model fits (also for linear models lm and glm) based on Wald statistics. However, it does neither support lme4::glmer for generalized linear model fits nor does it allow to estimate R2 for combinations of predictors. Furthermore, it does not estimate structure coefficients, inclusive R2 or part R2 for multilevel factors as a unit.

Features of partR2

partR2 takes a fitted (generalized) linear mixed-model (GLMM), from the popular mixed model package lme4 (Bates et al., 2015) and estimates part R2 by iteratively removing fixed effects (Nimon et al., 2008). The specific fixed effects of interest are specified by the partvars and/or by the partbatch argument. The package estimates part R2 for all predictors specified in partvars individually and in all possible combinations (the maximum level of combinations can be set by the max_level argument). A custom specification of fixed effects of interest saves computation time as compared to an all-subset specification and is therefore required in partR2.

The central function partR2 will work for Gaussian, Poisson and binomial GLMMs. Since the model fit is done externally, there is no need to supply a family argument. For non-Gaussian GLMMs, the package estimates link-scale R2 (sensu Nakagawa & Schielzeth, 2013). We implement parametric bootstrapping to quantify sampling variance and thus uncertainty in the estimates. Parametric bootstrapping works through repeated model fitting on simulated data based on fitted values (Faraway, 2015). The number of bootstrap iterations is controlled by the nboot argument. We recommend a low number of nboot for testing purposes and a large number (e.g., nboot = 1000) for the final analysis.

The package returns an object of class partR2 that contains elements for part R2, inclusive R2, structure coefficients, beta weights (standardized regression slopes), bootstrapping iterations and some other information. An extended summary, that includes inclusive R2, structure coefficients and beta weights can be viewed using the summary function. The forestplot function shows a graphical representation of the variance explained by individual predictors and sets of predictors along with their bootstrapping uncertainties. All computations can be parallelized across many cores based on the future and furrr packages (Vaughan & Dancho, 2018; Bengtsson, 2020). An extended vignette with details on the complete functionality accompanies the package.

Example with Gaussian data

We use an example dataset with hormone data collected from a population of captive guinea pigs to illustrate the features of partR2. The dataset contains testosterone measurements of 31 male guinea pigs, each measured at 5 time points (age between 120 and 240 days at 30-day intervals). We analyze log-transformed testosterone titers and fit male identity as a random effect. As covariates the dataset contains the time point of measurement and a rank index derived from behavioral observations around the time of measurement (Mutwill et al., 2021).

Rank and Time are correlated in the dataset (r = 0.40), since young individuals are typically low rank, while older individuals tend to hold a high rank. Time might be fitted as a continuous predictor or as a factor with five levels. Here we present the version of a factorial predictor to illustrate the estimation of part R2 for interactions terms. Hence, an interaction between Time and Rank will also be fitted.

First, the package needs to be loaded (after successful installation) in an R session (R Core Team, 2021). The package comes with the guinea pig dataset that also needs to be loaded using the data function.



A single record contains missing values for testosterone measurements. Missing records can be problematic to handle in partR2 and are better removed prior to the analysis. We also log-transform the response and convert Time to a factor and filter for the first three time points to simplify the output.

GuineaPigs <- subset(GuineaPigs, ! & ! & (Time %in% c(1,3,5)))

GuineaPigs$TestoTrans <- log(GuineaPigs$Testo)

GuineaPigs$Time <- factor(GuineaPigs$Time)

We then fit a linear mixed effects model using lmer from the lme4 package (Bates et al., 2015). Further exploration of the data and model checks are omitted here for simplicity, but are advisable in real data analysis.


mod <- lmer(TestoTrans ∼Rank * Time + (1—MaleID), data=GuineaPigs)

The partR2 analysis takes the lmer model fit (an merMod object) and a character vector partvars indicating the fixed effects to be evaluated. Interactions are specified with the colon syntax (see the package’s vignette for further details).

res <- partR2(mod, partvars = c(”Rank”, ”Time”, ”Rank:Time”), nboot=100)

The function returns a partR2 object. The print function reports the part coefficients of determination and a more extensive summary can be viewed with the summary function which also shows inclusive R2, structure coefficients and beta weights (standardized slopes) (Fig. 2).

Summary output for example data analysis with Gaussian data (guinea pig analysis).

Figure 2: Summary output for example data analysis with Gaussian data (guinea pig analysis).


summary(res, round_to = 2)

The variances appear largely additive, since combinations of predictors explain about the sum of the variance explained by individual predictors. The main components of the partR2 object can be accessed for further processing as res$R2 for part R2 (with point estimates and confidence intervals), res$SC for structure coefficients, res$IR2 for inclusive R2 and res$BW for beta weights.

Dealing with interactions

Models with interactions are problematic, because the variance explained by a main factor can be estimated in multiple ways (Fig. 3) and because of the internal parametrization of the model matrix.

Conceptual framework for dealing with interactions.

Figure 3: Conceptual framework for dealing with interactions.

An interaction is the product of two main effects and thus often correlated with each of the main effects. The figure shows three options for estimating the part R2 for main effects that are involved in an interaction.

The model output in Fig. 2 shows the number of parameters fitted in each model (each row in the R2 part refers to a reduced model). In the print and summary output this is visible as a column labelled ‘ndf’. A close inspection shows that the removal of rank did not change the number of parameters (6 for the full model, 6 for the model excluding rank). This is because the model matrix is reparametrized in the reduced model and lmer will fit three terms for the interaction (here Time1:Rank, Time3:Rank, Time5:Rank) rather than just two for the interaction in the full model. Dummy coding of the factor can be usefully combined with centering of dummy coded variables (Schielzeth, 2010) and gives more control over this re-parametrisation. It allows for example to estimate the part R2 for the average effect of Rank by constraining the average Rank effect to zero, so that only the two contrasts are fitted (here Time3:Rank, Time5:Rank):

GuineaPigs <- cbind(GuineaPigs, model.matrix(∼0 + Time, data=GuineaPigs))

GuineaPigs$Time3 <- GuineaPigs$Time3 - mean(GuineaPigs$Time3)

GuineaPigs$Time5 <- GuineaPigs$Time5 - mean(GuineaPigs$Time5)

The model can then be fitted with dummy predictors. Since the usual specification in partR2 via partvars would fit all possible combinations, including combinations of the different Time terms, such a run can take a long time. However we are mostly interested in fitting and removing all dummy predictors at a time. The package therefore features an additional argument partbatch to specify a list of character vectors containing the sets of predictors that should always be kept together. In the example, the list has two elements, a character vector for the dummy-coded main effects and a character vector for the interaction terms. The analysis yields part R2 for two batches of predictors as well as Rank and their combinations.

mod <- lmer(TestoTrans ∼(Time3 + Time5) * Rank + (1—MaleID),


batch <- c(”Time3”, ”Time5”)

partR2(mod, partvars=c(”Rank”), partbatch=list(Time=batch, “Time:Rank”= paste0(batch, ”:Rank”)), nboot=100)

This, however, is only one way of dealing with interactions (Option A in Fig. 3). It represents the variance uniquely explained by main effects even in the presence of an interaction. Since interactions are the products of main effects, interaction terms are typically correlated with main effects and the part R2 calculated above might not represent a biologically relevant quantity. There are two alternative ways of how to deal with interactions. Both are possible in partR2, but since requirements differ between applications, we do not implement one with priority.

One way to think about variance explained by main effects and their interactions is to pool the variance explained by a main effect with the variance explained by interactions that the term is involved in (Option B in Fig. 3). In the guinea pig example, for instance, Rank might be considered important either as a main effect or in interaction with time and we might want to estimate the total effect of Rank. This can be done for the guinea pig dataset by using partbatch:

mod <- lmer(Testo ∼Time * Rank + (1—MaleID), data=GuineaPigs)

partR2(mod, partbatch = list(Time=c(”Time”, ”Time:Rank”), Rank=c(”Rank”, ”Time:Rank”)), nboot=100)

A third, which we think usually preferable option is to prioritize main effects by assigning the proportion of variance that is explained by a main effect together with the variance jointly explained with its interaction to the main effect (Option C in Fig. 3). This implies that part R2 for a main effect is estimated when its own interaction is excluded from the model (mod1 and part1 below). The variance explained by the interaction is then estimated in a separate model (mod2 and part2 below). We have implemented a helper function mergeR2 that allows to merge two partR2 runs.

mod1 <- lmer(Testo ∼Time * Rank + (1—MaleID), data=GuineaPigs)

part1 <- partR2(mod1, partvars = c(”Time:Rank”), nboot=100)

mod2 <- lmer(Testo ∼Time + Rank + (1—MaleID), data=GuineaPigs)

part2 <- partR2(mod2, partvars = c(”Time”, ”Rank”), nboot=100)

mergeR2(part1, part2)

All these results can be viewed by print, summary and plotted by forestplot. It is important to bear in mind the differences in the interpretation as illustrated in Fig. 3.

An example with proportion data

As an example for proportion data, we analyze a dataset on spatial variation in color morph ratios in a color-polymorphic species of grasshopper. Individuals of this species occur either in a green or a brown color variant and the dataset contains counts of brown and green individuals (separated for females and males) from 42 sites sampled in the field (Dieker et al., 2018). Site identity will be fitted as a random effect. As covariates the dataset contains a range of Bioclim variable that describe various aspects of ecologically relevant climatic conditions (Karger et al., 2017). The aim is to identify the climatic conditions that favour one or the other colour variant.

We first load the grasshopper dataset. We standardise all Bioclim variables using the scale function and add an observation-level counter that will be used as an observation-level random effect (OLRE) to account for overdispersion (Harrison, 2014).


for (i in which(substr(colnames(Grasshoppers),1,3)==”Bio”)) {Grasshoppers[,i] <- scale(Grasshoppers[,i]) }

Grasshoppers$OLRE <- 1:nrow(Grasshoppers)

We first fit a GLMM with binomial error structure and logit link using the glmer function from the lme4 package (Bates et al., 2015). A previous analysis has shown that the first principle component of the Bioclim data explains a small, but significant part of variation in morph ratios (Dieker et al., 2018). For illustration, we use the four Bioclim variables that show a loading of more than 0.30 on the first principle component.

Comparison of part R2 for individual predictors (A), inclusive R2 (B), structure coefficients (C) and beta weights (D) for an example dataset with proportion data from grasshoppers.

Figure 4: Comparison of part R2 for individual predictors (A), inclusive R2 (B), structure coefficients (C) and beta weights (D) for an example dataset with proportion data from grasshoppers.

mod <- glmer(cbind(nGreen, nBrown) ∼  Bio7 + Bio14 + Bio17 + Bio19 +

(1—SiteID) + (1—OLRE), data=Grasshoppers, family=”binomial”)

res <- partR2(mod, partvars=c(”Bio7”, ”Bio14”, ”Bio17”, ”Bio19”),

max_level=1, nboot=100)

The summary output informs us (at the bottom) that there have been warnings in the bootstrapping processes. This is not unusual since bootstrapping frequently generates data, for which one of the parameters is estimated at the boundary (in particular if one of the variance components is very small). The results can be visualised using the forestplot function (Fig. 4). Plotting is based on ggplot2 (Wickham, 2016), and multiple forest plots can easily be assembled using the patchwork package (Pedersen, 2020). Forest plots show the effect sizes graphically and can be set to either show part R2 when type = ”R2” (the default), inclusive R2 when type = ”IR2”, structure coefficients when type = ”SC”, and beta weights (standardized model estimates) with type = ”BW”.

p1 <- forestplot(res, type = ”R2”)

p2 <- forestplot(res, type = ”IR2”)

p3 <- forestplot(res, type = ”SC”)

p4 <- forestplot(res, type = ”BW”)


(p1 + p2) / (p3 + p4) + plot_annotation(tag_levels = ”A”, tag_prefix = ”(”, tag_suffix = ”)”)

A comparison of part R2, inclusive R2, structure coefficients beta weights shows the different insights that can be gained from these different summaries of the model fit (Fig. 3). In this case, three of the Bioclim variables (Bio14, Bio17, Bio19) are highly positively correlated (r ≥ 0.93), while a fourth one (Bio7) is moderately negatively correlated to all three of them (r ≤  − 0.63). Part R2 are thus low, because none of the parameters uniquely explains a large share of the variance. Bio17 seems to be the best predictor of morph ratios, with the largest (negative) beta weight, largest part R2, largest structure coefficients and largest inclusive R2. Beta weights for the two positively correlated (but slightly weaker) predictors, Bio14 and Bio19, switch sign as is not unusual for collinear predictors. This means that after accounting for the effect of Bio17, they contribute positively to prediction. However, structure coefficients show that both variables load negatively on the linear predictor, as does Bio17.


Using transformation or functions in the formula argument can lead to issues with matching the terms of the model with the partvars argument of partR2. It is therefore important that the names in partvars match exactly the terms in the merMod object. However, any complications are easily circumvented by implementing the transformations before fitting the model and storing them in the data frame used in the analysis. It is also worth to be aware that unusual names may cause complications and renaming can offer an easy solution.

We have repeatedly seen model outputs where the point estimate does not fall within the confidence interval. This might seem like in the bug in the package, but in our experience usually indicates issues with the data and/or the model. In fact, parametric bootstrapping can be seen as a limited form of posterior predictive model checks (Gelman & Hill, 2006). If generating new data from the fitted model (as done with parametric bootstrapping) results in data that are dissimilar to the original data, then the model is probably not a good fit to the data.

Bootstrap iterations can sometimes yield slightly negative estimates of part R2, in particular if the variance explained by a predictor is low. These negative estimates happen in mixed-effects models, because estimates of random-effect variance might change when a predictor is removed and this can lead to a slight decrease in the residual variance, and hence a proportional increase in R2 (Rights & Sterba, 2019). By default, partR2 sets negative R2 values to 0, but this can be changed by setting allow_neg_r2 to TRUE. It also happens that inclusive R2 is estimated slightly lower than part R2 when the contribution of a particular predictor is very large. We consider both cases as sampling error that should serve as a reminder that variance components are estimated with relatively large uncertainly and minor differences should not be over-interpreted.

A warning needs to be added for the estimation of R2 (and, in fact, also repeatability R) from small datasets. In particular if the number of levels of random effect is low, variance components might be slightly overestimated (Xu, 2003). This issue applies similarly to the variance explained by fixed effects, in particular if the number of predictors is large relative to the number of data points.

Code and data availability

The current stable version of partR2 can be downloaded from CRAN ( and the development version can be obtained from GitHub ( The data used in the examples is part of the package.

Supplemental Information

Annotated example code for partR2 analysis

DOI: 10.7717/peerj.11414/supp-1
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