Fitting and evaluating univariate and multivariate models of within-lineage evolution

Abstract

The nature of phenotypic evolution within lineages is central to many unresolved questions in paleontology and evolutionary biology. Analyses of evolutionary time-series of ancestor-descendant populations in the fossil record are likely to make important contributions to many of these debates. However, the limited number of models that have been applied to these types of data may restrict our ability to interpret phenotypic evolution in the fossil record. Using uni- and multivariate models of trait evolution that make different assumptions regarding the dynamics of the adaptive landscape, I evaluate contrasting hypotheses to explain evolution of size in the radiolarian Eucyrtidium calvertense and armor in the stickleback Gaserosteus doryssus. Body size evolution in E. calvertense is best explained by a model where the lineage evolves as a consequence of a shift in the adaptive landscape that coincides with the initiation of neosympatry with its sister lineage.

Multivariate evolution of armor traits in a stickleback lineage (Gasterosteus doryssus) show evidence of adaptation towards independent optima on the adaptive landscape at the same time as traits change in a correlated fashion. The fitted models are available in a the R package evoTS, which builds on the commonly used paleoTS framework.

Introduction

Evolutionary time-series of populations in the fossil record provide information on phenotypic change on time intervals in-between generational and macroevolutionary time scales. Analyses of these types of data are thus poised to make important contributions to our current understanding of evolution across the time scale continuum. For more than 15 years, Hunt’s paleoTS framework (Hunt 2006; 2008a 2008b; Hunt et al. 2008; 2015; Hunt and Carrano 2010) has been instrumental in generating new knowledge of evolutionary trait dynamics at the intersection between paleontology and evolutionary biology analyzing such time-series. For example, trait evolution within lineages in the fossil record was shown to be much more diverse than only stasis (Hunt 2007; Hopkins and Lidgard 2012; Hunt et al. 2015), micro-evolutionary parameters have been successfully estimated from fossil data (Hunt et al. 2008), and rates of evolution can be estimated using similar models as in phylogenetic comparative methods (Hunt 2012). The new insights provided by the paleoTS framework on phenotypic evolution has thus contributed to a closer integration of paleontology and evolutionary biology.

Despite its success and impact, an extension of the paleoTS framework may be useful. First, a common use of the paleoTS R package is to investigate the relative fit of the three canonical models □—stasis, unbiased random walk and trend (modeled as an biased random walk) □—to fossil time-series. These models have a long history within paleontology (e.g., Raup 1977; Roopnarine 2001; Sheets and Mitchell 2001) but are not always able to adequately capture trait dynamics within lineages in the fossil record (Voje et al. 2018; Voje 2018). Fitting and comparing a larger range of evolutionary models may enable a richer interpretation of evolutionary change within lineages (Fig. 1).

Figure 1. Univariate evolution models that can be fitted and compared in evoTS.

The models stasis, strict stasis, biased and unbiased random walk and OU with fixed optimum are implemented in paleoTS (Hunt 2006; Hunt et al. 2008; 2015). The other models are implemented in evoTS. All models can be fitted and compared in evoTS. In the OU model with a moving optimum, the population is either displaced from the optimum at the start of the sequence or is residing on or very close to the optimum (latter model indicated by *). The dotted horizontal line shows the position of the optimum in the OU model with a fixed optimum and the starting value of the optimum for the model where the optimum is allowed to change.

Second, the adaptive landscape has been suggested as a conceptual bridge between our understanding of microevolutionary processes and evolution observed across longer time scales (Simpson 1944; Arnold et al. 2001; Hansen 2012). However, knowledge of the dynamics of the adaptive landscape across time is poor. Macroevolution is likely associated with movements of peaks on the adaptive landscape, but a fixed adaptive landscape is commonly assumed in microevolutionary studies, which is also the case for the models implemented in paleoTS. Inferring the dynamics of the adaptive landscape from evolutionary time-series may contribute towards a better understanding of the dynamical nature of peak movements at different time intervals.

Third, evolution is inherently a multivariate phenomenon. Pleiotropy is omnipresent (e.g., Walsh and Blows 2009) and selection on one trait may cause genetically linked traits to evolve (Lande 1979; Lande and Arnold 1993). Traits may also commonly experience correlated selection. Multivariate models are useful for investigating whether traits change in a correlated or uncorrelated manner, whether one trait/variable affects the optimum of a second trait, or whether adaptation in traits happen independently of each other. The univariate models in paleoTS are of limited use for assessing the consequences of the multivariate nature of selection and evolution within lineages.

Here, I explore these three avenues of research by fitting uni- and multivariate models to examine evolution of size in a radiolarian lineage and multivariate evolution of armor traits in a stickleback lineage. Analyses employ the new R package, evoTS. As the univariate models in evoTS are natural extensions and modifications of the models in paleoTS, I start by introducing the univariate models available in paleoTS before I explain the expanded univariate models implemented in evoTS. I then apply these models to a well-known and previously published data set, the evolution of size in the radiolarian lineage Eucyrtidium calvertense during allopatry and in a subsequent phase of neosympatry with its sister lineage E. matuyamai (Kellogg 1975). I continue by introducing the multivariate models implemented in evoTS before I apply them in a reanalysis of a published data set of two armor traits in a stickleback lineage (Bell et al. 2006; Hunt et al. 2008).

evoTS is Compatible with paleoTS

I have developed evoTS to mirror the user experience from paleoTS as much as possible. The two frameworks use the same data format and the model fitting procedures are built on the same assumptions. For example, all models assume the population (sample) means in the sequence of ancestor-descendants have a joint distribution that is multivariate normal with an expected mean vector and covariance matrix that are functions of the parameters of each model, the time intervals separating the populations (samples) in the sequence, and the sampling variances of the trait means calculated for each population (sample). The expected distribution of sample means is thus defined by their means, variances and covariances given the assumption of multivariate normality. All models in evoTS have been implemented using the joint parameterization routine from paleoTS (Hunt 2008a), with the autocorrelation among samples being accounted for in the log-likelihood function. As in the paleoTS package, evoTS uses a quasi-Newton optimization routine for estimating maximum likelihood parameter estimates for univariate models, while the Nelder-Mead hill climbing algorithm is the default option for some of the multivariate models. Relative model fit is evaluated based on the small sample-corrected version of the AICc (Akaike 1974; Burnham and Anderson 2002).

Univariate Models in paleoTS

Unbiased random walk, biased random walk (trend model), and stasis were the first models implemented in the paleoTS framework (Hunt 2006). An unbiased random walk models evolution of a trait mean as independent draws from a normal distribution with mean zero (μ = 0) and a variance $\left({\sigma }_{step}^2\right)$ commonly referred to as the step variance (Hunt 2006). Each draw represents a discrete evolutionary “step” and the expected amount of evolution in the trait mean z per time step i is 0

$$E\left[z_i\right]=z_0.$$ (1)

The trait mean is therefore not expected to be different from the ancestral state z_o, but the variance around this expectation increases linearly with elapsed time,

$$Var\left[z_i\right]={\sigma }_{step}^2{t}_i+{\epsilon }_i,$$ (2)

where t_i is the elapsed time from the start of the time-series to sample i (time at the start of the time-series is always 0). The variance in each sample is influenced by the sampling error (ε_i) in estimating the trait mean, which is equal to the sample variance divided by the sample size (i.e., number of measured specimens) for that sample. The covariance among sample means is given by

$$Cov\left[Z_i,Z_j\right]={\sigma }_{step}^2{t}_{\min ,}$$ (3)

where t_min represents the time interval between the start of the sequence and the earliest of the samples z_i and z_j.

The biased random walk (sometimes referred to as a trend model, see e.g., Hansen (1997)) is identical to an unbiased random walk except for a non-zero mean (μ ≠ 0) of the normal distribution from which evolutionary steps are drawn. A larger deviation from 0 of the mean translates into a stronger tendency to change unidirectionally in trait space. The expressions for the variance and covariance are identical for the biased and unbiased random walk (eq. 2 and 3), while the expected mean trait value is given by

$$E\left[z_i\right]=z_o+\mu t_i.$$ (4)

Various definitions of stasis have been employed in research that aims to quantify change in evolutionary time-series (e.g., Gingerich 1993; Bookstein 1987; Roopnarine 2001; Sheets and Mitchell 2001). The stasis model in paleoTS is similar to a white noise process where trait evolution consists of uncorrelated fluctuations around a fixed trait value (θ) (Hunt 2006). The fluctuations around the fixed mean are described by a variance parameter (ω), which is assumed to stay constant over time. Time is accordingly not a relevant parameter in the stasis model. The strict stasis model (Hunt et al. 2015) is identical to the previously described stasis model, except that ω = 0, which can be the case when the variance among trait means is smaller than the sampling error in the trait means, i.e., the observed differences among trait means can be explained by sampling error alone (see also Hannisdal 2006).

Hunt et al. (2008) extended the paleoTS framework with the implementation of an Ornstein-Uhlebeck (OU) model describing evolution of a trait towards a fixed peak in the adaptive landscape (Hansen 1997). The OU process is the simplest stochastic model that allows evolution toward a specific state and is given by the following differential equation

$$dy=\alpha (\theta -y)dt+{\sigma }_{y}d{W}_y,$$ (5)

where dy is the change in the trait (y) over a time step dt, α describes the rate of evolution towards the optimum θ, dW_y represents independent and normally distributed changes with mean 0 and unit variance, with σ_y being the standard deviation of this white-noise process. The first part of the OU process is deterministic and describes how the trait is pulled towards the optimum at a rate given by α. The second part is a stochastic process adding random noise scaled by the σ_y parameter to the trait dynamics. The expected change in a trait mean z per time step i and its variance and covariance are given by

$$E\left[z_i\right]=e^{\left(-\alpha t_i\right)}{z}_o+\left(1-e^{\left(-\alpha t_i\right)}\right)\theta ,$$ (6)

$$Var\left[z_i\right]=\left({\sigma }_{step}^2/2\alpha \right)\left[1-e^{\left(-2\alpha t_i\right)}\right]+{\epsilon }_i,$$ (7)

$$Cov\left[z_i,z_j\right]=\left({\sigma }_{step}^2/2\alpha \right)e^{\left(-\alpha t_{ij}\right)}\left[1-e^{\left(-2\alpha t_{\text{min}}\right)}\right],$$ (8)

where θ is the optimal trait value and t_ij is the time separating samples i and j (Hansen 1997; Hunt et al. 2008).

Univariate Models in evoTS

Simple models often sacrifice precision and nuance to distill general properties from data (Levins 1996). An evolutionary time-series showing a relative better fit to an unbiased random walk compared to a stasis model does not mean trait evolution was random in each generation in the analyzed lineage. Rather, it suggests that the observed trait dynamics is more consistent with a pattern of “meandering” evolution, in which random changes in the trait mean accumulate over time, rather than with random fluctuations around a constant mean (akin to a white noise process). Adding or changing a few parameters in the models implemented in paleoTS can aid in extracting additional information not captured by the original models.

Below, I describe the implemented univariate models in evoTS and briefly discuss how they can be interpreted when fitted to evolutionary sequences. evoTS is available for download from the Comprehensive R Archive Network (CRAN) (http://cran.r-project.org/). The online vignette contains a detailed walk-through that explains from a user perspective how to fit all the different univariate models in evoTS (and paleoTS) and how to evaluate their relative fit to data (klvoje.github.io/evoTS/index.html).

Time-Varying Unbiased Random Walks

The rate of evolution is constant in an unbiased random walk, which means the trait variance is expected to increase linearly with time. A natural extension of this model is to allow the rate of evolution to change with time. The decelerated model of evolution implemented in evoTS is an unbiased random walk where the step variance is reduced exponentially through time (Voje 2020). This model is basically identical to the early burst model developed for phylogenetic comparative data to test for a decelerated rate of evolution at the clade level (e.g., Cooper and Purvis 2010; Harmon et al. 2010). The expected evolutionary divergence between ancestor and descendant populations is zero in the decelerated evolution-model (eq. 1) and its variance and covariance are given by:

$$Var\left[z_i\right]={\sigma }_{step.0}^2{e}^{r{t}_i}+{\epsilon }_i,$$ (9)

$$Cov\left[z_i,z_j\right]={\sigma }_{step\text{.}0}^2{e}^{r{t}_{\min }},$$ (10)

where σ²_step.0 is the initial value for the step distribution, and r describes the exponential decay in σ²_step.0 through time and is thus constrained to be < 0. An accelerating model of evolution is identical to the decelerated model, except that the r parameter is constrained to be > 0. The time it takes to half (for the decelerated evolution model) or double (for the accelerated evolution model) the rate of evolution is given by ln(2)/|r|. The estimating algorithm in evoTS generally produces precise estimates of the r parameter in the decelerated and accelerated models (more details are given in Supplementary material Figure 1, see also Voje 2020).

Model Interpretation

A linearly increasing trait divergence with time will rapidly produce magnitudes of evolutionary change rarely observed out in nature (e.g., Lynch 1990; Gingerich 2001; Estes and Arnold 2007; Uyeda et al. 2011). A decelerating rate of evolution mitigates this problem. Although many evolutionary scenarios and processes can be compatible with unbiased random walk models, a decrease in the rate of evolution over time might for example occur when a lineage experiences less selection after a period of higher initial rates of evolution due to changes in the environmental conditions. (e.g., Voje 2020). A reduced rate may also occur if the effect of drift is reduced over time (i.e., due to increasing population size). An accelerated rate of evolution is not sustainable across long time scales but might fit lineages that experience an increased effect of drift or experience increasing environmental perturbations. Note that interpreting the trait dynamics as the results of neutral drift only may not be a plausible interpretation for ecologically relevant traits across long time scales (e.g., Hansen 2012). An alternative – and perhaps more likely – interpretation of unbiased random walk models is that they provide information on peak movements of the adaptive landscape. This is indeed a common interpretation of the related Brownian motion model in phylogenetic comparative approaches (Felsenstein 1988). As long as populations are sufficiently evolvable to immediately track changes in the locations of peaks in the adaptive landscape, the step variance $\left({\sigma }_{step}^2\right)$ provides insight on the rate of change of the adaptive peak itself according to this interpretation of the model. The decelerated and accelerated models may therefore represent scenarios where the rate of peak movements changes with time.

Ornstein-Uhlenbeck Models with Moving Optimum

A natural extension of the fixed peak OU model implemented in paleoTS is to allow the peak to change. A model where the optimum is changing according to a Brownian motion was proposed by Hansen et al. (2008) for analysis of phylogenetic comparative data. Adjusted to describe evolution of a single lineage, the expected trait mean is given by equation 5, while the variance and covariance are given by the following expressions:

$$Var\left[z_i\right]=\left[\frac{{\sigma }_{step}^2+{\sigma }_{\theta }^2}{2\alpha }\right]\left(1-e^{\left(-2\alpha t_i\right)}\right)+{\sigma }_{\theta }^2{t}_i\left[1-2\left(1-e^{-\alpha t_i}\right)/\alpha t_i\right]+{\epsilon }_i$$ (11)

$$Cov\left[z_i,z_j\right]=\left[\frac{{\sigma }_{step}^2+{\sigma }_{\theta }^2}{2\alpha }\right]\left(1-e^{\left(-2\alpha t_a\right)}\right)e^{-\alpha t_{ij}}+{\sigma }_{\theta }^2{t}_a\left[1-\left(1+e^{-\alpha t_{ij}}\right)\left(1-e^{-\alpha t_a}\right)/\alpha t_i\right]$$ (12)

where θ₀ is the initial (ancestral) optimum, σ²_θ is the (step) variance of the stochastic perturbations of the optimum, and t_α is the time interval from the ancestral population to the oldest of the two populations z_i and z_j. The half-life, ln(2)/α, is a reparameterization of the speed of adaptation in this process that is easy to interpret, as it is the time it takes for the trait to move half-way from the ancestral state to the optimum. The estimation algorithm is able to identify model parameters well, but outliers occur. Precision increases with longer time-series (see supplementary material Figure 2 for more details).

Model Interpretations

The stability of the adaptive landscape is debated and is likely affected by many factors (e.g., Slater and Friscia 2019). A lineage in a hyper-stable niche may reside on a fixed peak, while a lineage inhabiting a more unstable environment may experience a more dynamic adaptive landscape. For example, traits with specialized ecological roles insensitive to changes in overall size (i.e., allometry) may reside on stable peaks, while the peak of a size-associated trait easily affected by changing ecological conditions may be in constant flux. Being able to explicitly test whether a fixed or a dynamic optimum model best fit a given evolutionary sequence may provide a valuable perspective on the dynamical nature of the adaptive landscape.

Mode-Shift Models

There is no a priori reason why a lineage should be described by a single evolutionary process (Hunt 2008b, Hunt et al. 2015). Mode-shift models allow two or more separate segments of a time-series to evolve according to different models. evoTS contains a function that allows testing all pairwise combinations of the models unbiased random walk, biased random walk, stasis and OU, including independently parameterizing the same model to two separate segments. In addition to assessing all possible switch points in mode of evolution, it is also possible to define where in the sequence a shift in mode occurs, a helpful feature if we have an a priori hypothesis for when a shift happened.

Applying the Univariate Models

Changes in the adaptive landscape may affect how lineages evolve. I reinvestigate an evolutionary sequence of a radiolarian lineage to assess the dynamics of the adaptive landscape across a few million years and how it affects size change in the lineage.

Background

Kellogg (1975) investigated whether size evolution in the radiolarian lineage Eucyrtidium calvertense showed trait dynamics consistent with a scenario of character displacement (Fig. 2). E. matuyamai evolved from E. calvertense in subarctic waters and the two lineages differentiated during a period of allopatry. The two species came into secondary contact when a population of E. matuyamai migrated to sub-tropical waters. During this neosympatric phase, the two lineages differentiated in size, with E. mutuyamai evolving to become about 25% larger and E. calvertense to become about 10% smaller. Kellogg (1975) concluded that the evolutionary sequence of E. calvertense in sub-tropical waters showed little net change during the allopatric phase, but a trend towards smaller size in the neosympatric phase, a type of trait dynamics Kellogg (1975) interpreted to be consistent with the process of character displacement. The evolutionary sequence spans 3.67 million years and consists of 49 samples with a median and mean number of measured specimens per sample of 25 and 25.4, respectively (Fig. 2). The allopatric and neosympatric phases last for about 1.70 and 1.97 million years.

Figure 2. Size evolution in E. calvertense (Kellog 1975).

The vertical grey bar indicates the shift from allopatry to sympatry with E. mutuyamai. Blue dots belong to the allopatric phase and orange points belong to the sympatric phase. The best model is a mode-shift model consisting of two OU processes with fixed optima. The maximum likelihood parameter estimates (± standard error) of this model are: z_o = 4.543 (±0.019), θ₁ = 4.524 (±0.009), θ₂= 4.377 (±0.021), ${\sigma }_{step.1}^2=0.183(\pm 0.130)$, ${\sigma }_{step.2}^2=0.046(\pm 0.027)$, α₁ = 94.282 (±64.671), α₂ = 18.833 (±10.231). The broken horizontal lines represent the fixed optimal trait values from the OU-OU model.

Fitted Models

A mode-shift model consisting of two OU models (i.e., an OU-OU model) can assess if the initiation of the neosympatric phase lead to a sudden change in the position of the adaptive peak for size in E. calvertense. I also fitted OU processes with a changing optimum to investigate how well models assuming continuous change of the adaptive landscape explained the data. To investigate whether models assuming a fixed adaptive landscape outcompeted the models assuming a dynamic landscape, I fitted the stasis model, the trend model (i.e., a biased random walk), and unbiased random walk model with fixed, decelerating and accelerating rates of evolution, and mode shift models where the allopatric and neosympatric parts were either modeled as two unbiased random walks or where the second model was a biased random walk. Data and R scripts for replicating the analyses are available in the Supplementary Material.

Results

The OU-OU mode-shift model showed the best relative fit to the data (Table 1), with an optimal trait value during the sympatric phase (4.38 log micrometer) that was 13% smaller compared to the optimum during the sympatric phase (4.52 log micrometer). The adaptive process is faster in the allopatric phase (ln(2)/α = 0.007, which translates into a half-life of about 27 000 years) compared to the sympatric phase (ln(2)/α = 0.037, i.e., a half-life of about 135 000 years, Table 2). The log-likelihood surfaces of the half-life values show some overlap in the two phases (Fig. 3), but while a half-life of 4.05% (about 150 000 years) of the sequence length is within 2 log-likelihood units in the allopatric phase, the equivalent value of the second phase is 31.51% (about 1 160 000 years). The stochastic part of the trait dynamics is also elevated during the allopatric phase $({\sigma }_{step\text{.}1}^2=0.183)$ compared to the sympatric phase $\left({\sigma }_{step.2}^2=0.046\right)$. To investigate if the difference in temporal resolution between the two segments could explain the difference in trait dynamics, I sub-sampled the first segment 1000 times to match the length of the second segment (14 samples) and re-estimated the half-life and ${\sigma }_{step}^2$ parameters. The median estimates of the half-life and ${\sigma }_{step}^2$ from the sub-sampled data were 0.010 and 0.111, which suggests that differences in temporal resolution alone cannot explain the difference in the estimated trait dynamics between the two segments. The allopatric phase therefore appears to be characterized by faster evolution towards the fixed optimum and larger stochastic deviations from the optimum, compared to the sympatric phase.

Table 1.

Model fit to the E. calvertense sequence. The log-likelihood (log-lik.) and the relative model fit for the candidate models fitted to the evolutionary sequence of E. calvertense.

Model type	Model	K	log-lik.	AICc	delta AIC	AICc weight
No mode shift	Stasis	2	59.579	-114.896	55.562	0.000
	Biased random walk (trend)	3	87.422	-168.312	2.147	0.103
	Unbiased random walk	2	86.952	-169.644	0.815	0.201
	Decelerated evolution	3	86.657	-166.780	3.678	0.048
	Accelerated evolution	3	83.656	-160.778	9.680	0.002
	OU with fixed optimum	4	88.078	-167.247	3.211	0.061
	OU with moving optimum	4	89.478	-170.046	0.412	0.246
	OU with moving optimum*	5	88.078	-164.761	5.697	0.018
Mode shift	Two unbiased random walks	4	84.942	-160.974	9.484	0.003
	Unbiased and biased random walk	5	85.562	-159.729	10.729	0.001
	Two OU models	8	95.029	-170.458	0.000	0.303

Table 2.

Maximum likelihood parameter estimates for the candidate models fitted to the E. calvertense sequence. See equations and the main text for definitions of the different model parameters. The numbers in parentheses are standard errors calculated from the square-root of the inverse of the diagonal of the Hessian matrix.

Model	Parameter estimates (standard error)
Stasis	θ = 4.49 (0.01), ω = 0.01 (0.00)
Biased random walk	Zo = 4.54 (0.02), μ = -0.19 (0.19), ${\sigma }_{step}^2=0.04(0.05)$
Unbiased random walk	Zo = 4.54 (0.02), ${\sigma }_{step}^2=0.04(0.00)$
Decellerated evolution	Zo = 4.36 (0.06), ${\sigma }_{step}^2=0.05(0.00)$, r = -2.80 (1.14),
Accelerated evolution	Zo = 4.54 (0.23), ${\sigma }_{step}^2=0.05(\text{NaN})$, r = 0.54 (0.17),
OU with fixed optimum	Zo = 4.55 (0.02), ${\sigma }_{step}^2=0.04(0.02)$, θ = 4.40 (0.09), α = 3.21 (3.09
OU with moving optimum	Zo = 4.54 (0.02), ${\sigma }_{step}^2=0.22(0.29)$, ${\sigma }_{step.opt}^2=0.02(0.01)$, α = 205.57 (325.00)
OU with moving optimum*	Zo = 4.55 (0.02), θ0 = 4.40 (0.09), ${\sigma }_{step}^2=0.04(0.02)$, ${\sigma }_{step.opt}^2=0.00(-)$, α = 3.21 (2.70)
Two unbiased random walks	Zo = 4.52 (0.02), ${\sigma }_{step.1}^2=0.06(0.03)$, ${\sigma }_{step.2}^2=0.04(0.02)$
Unb. and biased random walk	Zo = 4.52 (0.01), ${\sigma }_{step\text{.}1}^2=0.06(0.03)$, μ= -0.21 (0.28), ${\sigma }_{step.2}^2=0.04(0.02)$
Two OU models	Zo = 4.54 (0.02), ${\sigma }_{step.1}^2=0.18(0.13)$, θ1 = 4.52 (0.01), α1 = 94.28 (64.67) ${\sigma }_{step\text{.2}}^2=0.05(0.03)$, θ2 = 4.37 (0.02), α2 = 18.83 (10.23)

Figure 3. Log-likelihood surfaces for the OU-OU model.

The panels show the support surface for the OU model describing the evolutionary sequence before and after the mode shift, respectively. The elevated area represents parameter estimates that are within two loglikelihood units of the best estimate. Panel A represents the first part of the sequence and the 2-unit support surface includes immediate adaptation (i.e., half-life = 0) and extends up to 0.040. Panel B represents the second part of the sequence where a half-life of zero is not part of the support surface (0.019 – 0.315). The range of support for the two stationary variances are 0.000 – 0.002 and 0.001 – 0.008). Note that these results are conditional on the best estimate of the other parameters in the model (i.e., the ancestral state and the optimum).

Models that differ in their relative model fit by a few AICc units may be worth considering as plausible or suitable alternative explanation of an empirical data set (Burnham et al. 2011). The OU model with a changing optimum has a very similar fit compared to the best model (ΔAICc = 0.412). The alpha parameter describing the rate of evolution towards the moving optimum is large (205.57), translating into a point estimate of the half-life of about 10,000 years. The point estimate of the rate of change in the optimum $\left({\sigma }_{step.opt}^2=0.02\right)$ clearly indicates a non-fixed optimum through time. The stochastic part of the trait dynamics is rather large $\left({\sigma }_{step}^2=0.22\right)$, suggesting size evolution in E. calvertense has contributions from both the deterministic and stochastic part of the OU model. A reasonable interpretation of the trait dynamics in E. calvertense according to the parameter values of this OU model is as a white noise process around a stochastically moving peak. Note that the unbiased random walk shows a similar, albeit somewhat poorer, fit to the data compared to the OU process with a moving optimum. This is not surprising since the optimum in the OU process changes according to an unbiased random walk. The better fit of the OU model is due to the size of the fluctuations around the optimum, which is sufficiently large not to be accounted for by measurement error in the samples. Not controlling for error in the samples would therefore unduly favor the unbiased random walk instead of the OU process.

In summary, the best model among the candidates suggests the position of the optimum changed towards a smaller optimal size when E. calvertense comes into secondary contact with E. mutuyamai. Evolution towards a randomly changing optimum in both the allopatric and sympatric phases of the evolutionary sequence, or an unbiased random walk, are also likely models of the trait dynamics in E. calvertense.

Multivariate Models in evoTS

Much can be learned from studies of single traits, but a trait-by-trait approach has some important shortcomings. The omni-presence of pleiotropy suggests there is only a very small number of truly genetically independent traits (Barton 1990; Johnson and Barton 2005; Walsh and Blows 2009). Evolutionary change in a trait is only rarely due to selection operating on that trait alone as selection on genetically linked traits may also affect the focal trait (Lande 1979; Lande and Arnold 1983; Hansen and Houle 2008). Traits that are genetically independent may still be functionally dependent, which means they may experience coordinated selection and therefore have a tendency to evolve in concert. Trait evolution is thus inherently a multivariate process that requires multivariate models to be more fully understood.

Below follows a description of the multivariate models available in evoTS and how they can be interpreted. The online vignette details from a user perspective how to fit the different multivariate models implemented in evoTS, including walk-troughs and examples of how to test different hypotheses of evolution and adaptation (klvoje.github.io/evoTS/index.html).

Multivariate Unbiased Random Walks

The multivariate unbiased random walk model can assess whether a set of traits evolve in a coordinated fashion or not. This is done by estimating an evolutionary rate matrix R (Felsenstein 1988; Revell and Harmon 2008). The R matrix describes the rate of evolution in the investigated traits on the diagonal (i.e., the diagonal contains the step variances) and the covariance of the changes in the traits in the off-diagonal elements. The multivariate variance-covariance matrix for the unbiased random walk model (V) is computed using the Kronecker product of the R matrix and a “distance matrix” C, describing how the different samples/populations are separated in time

$$\mathit{V}={\displaystyle \sum _{i=1}^m{\mathit{R}}_{\mathit{i}}\otimes {\mathit{C}}_{\mathit{i}},}$$ (13)

where m represents the number of non-overlapping segments of a time-series that have their own R matrix. Sampling error of the trait mean (calculated as the sample variance divided by the sample size) is added to the diagonal of V. To ensure symmetric positive definiteness of the V matrix during log-likelihood optimization, R is parameterized by its Cholesky decomposition as the cross-product of upper triangular matrices

$$\mathit{R}=\mathit{L}{\mathit{L}}^{\mathit{T}}$$ (14)

where L is a square matrix with positive diagonal entries. L is upper triangular if there are off-diagonal elements in R. As for the univariate unbiased random walk, it is possible to test for a decrease or increase in the rate of change over time in the multivariate unbiased random walk model in evoTS. The r parameter adjusting the rate is assumed common for all the traits. Simulations show that the estimating procedure produce unbiased parameters even at sequence lengths of about 10 samples (see Supplementary material Figure 3 for more info).

Model interpretation

Potential causes of correlated trait evolution are many. Traits may for example independently follow optima governed by the same environmental drivers, show concerted evolution due to shared direct or indirect selection, or be affected similarly by genetic drift.

Note that the R matrix is not the same as the genetic (G) or phenotypic (P) co-variance matrices commonly estimated in quantitative genetics. However, the R matrix is connected to these matrices under certain assumptions. For example, the R matrix is expected to be proportional to the additive genetic variance–covariance matrix (G) if the traits evolve due to genetic drift only (Lande 1979; Felsenstein 1988). Estimating R can thus aid in assessing to what extent evolution within lineages match predictions from quantitative genetics.

Multivariate Ornstein-Uhlenbeck Models

Multivariate evolution is more than correlated change. Multivariate versions of the OU-process allow for sophisticated investigations of a range of hypotheses regarding evolution and adaptation and is described by the following differential equation (Bartoszek et al. 2012; Reitan et al. 2012; Clavel et al. 2015):

$$dZ=\mathit{A}(\theta (t)-Z(t))dt+\mathit{R}dW(t)$$ (15)

where A is a square matrix (with dimensions equal to the number of traits) describing the rate of evolution toward the optimal trait values, θ is a vector containing the optimum for each trait, R is a square matrix (with dimensions equal to the number of investigated traits) describing the stochastic changes in the traits, and W is the diffusion parameter. Under the assumption that we only have one selective regime (optimum) per trait, the expected trait means of the Ornstein-Uhlenbeck process are the weighted sum of the optimum and the ancestral trait value (Hansen 1997):

$$E\left[Z_i(t)\right]=e^{\left(-A{t}_i\right)}{Z}_0+\left(1-e^{\left(-A{t}_i\right)}\right)\theta$$ (16)

where Z_i is a vector containing the expected trait values for sample i, z₀ is a vector containing the ancestral trait means, and t_i is the time interval from the ancestral population mean (the start of the time-series, which has a time of 0) to the ith population mean.

The variance and covariance of sample/population means are given by the following expression (Bartoszek et al. 2012; Reitan et al. 2012; Clavel et al. 2015):

$$Cov\left(z_i,z_j\right)=\left[Q\left({\left[\frac{1}{{\lambda }_k+{\lambda }_l}\left(1-e^{-\left({\lambda }_k+{\lambda }_l\right)t_a}\right)\right]}_{1\le k,l\le m}\odot Q^{-1}L{L}^T{\left(Q^{-1}\right)}^T\right)Q^T\right]e^{-A^T{t}_{ij}}$$ (17)

where Q is the orthogonal matrix of eigenvectors of A, LL^T is the Cholesky decomposition of the R matrix (R = LL^T), λ_i is the ith eigenvalue of A, t_a is the time interval from the ancestral population to the oldest of the two populations z_i and z_j, t_ij is the time separating two samples z_i and z_j, and ⊙ represents the Hadamard (element-wise) matrix product. Under the assumption that the A matrix has a number of linear independent eigenvectors equal to the number of traits investigated, A can be expressed by its eigendecomposition

$$A=Q\wedge Q^{-1}$$ (18)

where ⋀ is a matrix with the eigenvalues (λ) of A on the diagonal. The matrix exponential in equation 17 can then be solved using the eigendecomposition of A (Bartoszek et al. 2012; Reitan et al. 2012; Clavel et al. 2015):

$$e^{-At}=\mathit{Q}diag\left(e^{-{\lambda }_{1}t},\dots ,e^{-{\lambda }_{m}t}\right){\mathit{Q}}^{-1}$$ (19)

Estimating parameters precisely is a challenge that tends to increase with model complexity. Parameter estimation under univariate and multivariate phylogenetic OU models can be difficult (Hansen et al. 2008; Bartoszek et al. 2012; 2023; Ho and Ané 2014; Cressler et al. 2015), but simulations show that multivariate OU model parameters in evoTS are overall identifiable. Precision is high for diagonal elements in R and A and the optima already for short time-series. The median parameter estimate of the off-diagonal elements in A are in the proximity of the true value for short sequence lengths and approaches the true value with increasing sequence lengths (see Supplementary material Figure 4 for more info).

Model interpretation

The A and R matrices are key to define the trait dynamics in a multivariate OU model. The elements in R control the stochastic parts of the trait dynamics in the OU process and can be interpreted similarly as R in the multivariate unbiased random walk: The diagonal elements in R represent the step distributions (step variances) of the changes in each individual trait while any non-zero off-diagonal elements represent the covariance of the stochastic changes in the traits. The elements in the A matrix affect the deterministic part of the OU process, i.e., the adaptive process of traits evolving toward optima. The diagonal elements in A are the individual alpha parameters for each trait, while a non-zero off-diagonal element reflects how changes in the trait affects the optimum of another trait. Four broad categories of hypotheses can be investigated based on how the A and R matrices are parameterized.

Independent Trait Evolution

Both the deterministic and stochastic parts of the evolution are independent for each trait. This model is equivalent of fitting univariate OU models to each trait separately and is parameterized in the multivariate version by allowing only diagonal elements in the A and R matrices to be non-zero.

Independent Adaptation

Each trait adapts independently to their optimum, but the stochastic part of the trait dynamics is correlated. This model is obtained if the A matrix is diagonal while the R matrix has non-zero off-diagonal elements.

Non-independent Adaptation

Changes in trait X affect the optimum of trait Y, but changes in Y are not affecting the optimum of trait X. A can be a non-symmetric matrix (contrary to the R matrix), which means one trait is allowed to affect the optimum of another trait, but not vice versa. A negative number in an off-diagonal element in A means the trait is evolving towards the optimum while a positive number means the trait is repelled away from the optimum. The stochastic changes in the trait (controlled by the parameterization of the R matrix) can be either correlated or uncorrelated.

Reciprocal Adaptation

Traits affect each other’s optima. This is the case if their respective off-diagonal elements in A are non-zero. One trait may assert a larger effect on the optimum of another trait than vice versa. The stochastic changes in the trait can be either correlated or non-correlated.

An A matrix with non-zero off-diagonal elements investigates Granger causality between the two traits/variables (Granger 1969; Schweder 1970; Reitan et al. 2012; Hannisdal and Liow 2018; Reitan and Liow 2019). Granger causality is a statistical concept that is used to determine whether one time-series is useful in predicting another. Simply speaking, we have evidence for Granger causality if observations in one time-series is useful for forecasting future observations in one or several other time-series. The multivariate OU process therefore allows us to move beyond interpreting correlations among variables in time-series. A correlation is symmetric (the correlation between X and Y is the same as the correlation between Y and X), but that is not needed for Granger causality. X can (Granger) cause Y, without Y Granger-causing X. Forecasting using Granger causality is only possible if there is some lag in the tracking of the optimum i.e., if a trait is unable to immediately respond to changes in the optimum. The rate of adaptation in the multivariate OU model is determined by the entries in the A matrix and can be conveniently calculated by the half-life. Similarly as in the univariate OU model, the half-life (ln(2)/ α_ii) describes the time it takes for the trait to move half-way from the ancestral state to the new optimal state.

The multivariate OU model is reduced to an unbiased random walk if the diagonal elements in A are zero. Within the multivariate OU process, it is therefore possible to allow traits/variables that change according to an unbiased random walk to affect the optimum of traits evolving according to an OU process. Models with a mix of traits evolving as either OU or as an unbiased random walk has been named OUBM models in the phylogenetic comparative literature (e.g., Bartoszek et al. 2012; 2023). An OUBM model may for example be a sensible choice if we want to investigate if an environmental variable (e.g., a paleoclimatic proxy) that we assume evolves as an unbiased random walk affects the optimum of a trait we assume evolve according to an OU process.

Applying the Multivariate Models

Analyzing several traits or variables jointly using multivariate models enable a more sophisticated assessment of alternative hypotheses of the observed trait dynamics in the data relative to analyzing each trait separately. I apply multivariate models on a data set on armor trait evolution in a threespine stickleback lineage (Gasterosteus doryssus) which has previously been analyzed using univariate models (Bell et al. 2006; Hunt et al. 2008). Strong genetic covariances among armor traits (Leinonen et al. 2011) and evidence that different armor traits are affected by the same loci (Cresko et al. 2004) in extant threespine sticklebacks (Gasterosteus aculeatus) suggest armor traits may not evolve independently. Armor traits are also likely to serve similar ecological functions, and may therefore often experience similar selection pressures. A multivariate approach is therefore warranted when analyzing armor traits in threespine sticklebacks.

Background

Bell et al. (2006) analyzed morphological evolution in three skeletal traits in a fossil stickleback lineage (Gasterosteus doryssus) across more than 7000 years from well-preserved lake sediments. The three traits are part of the armor of the fish; number of dorsal spines, number of touching pteryigiophores and a pelvic trait measured based on scores of the completeness of the pelvic condition. All three traits show a clear trend of reduction in number (and score) during the time-series, likely as a consequence of a reduced predation pressure in the lake (Bell et al. 2006). However, Bell et al. (2006) did not find strong evidence for natural selection in these timeseries, as falsifying a null-model of neutral evolution (drift) proved difficult. Hunt et al. (2008) reanalyzed the data using paleoTS and found that an OU model with a fixed optimum showed a much better relative fit to each of the three stickleback traits compared to an unbiased random walk (drift model), thus supporting adaptive evolution towards trait-specific optima. The multivariate analysis of the data in this paper assess to what extent there is evidence for correlated and/or adaptive coevolution of the armor traits.

Respecting the scale type of the investigated variables in quantitative analyses is important to producing meaningful results (Houle et al. 2011; Voje et al. 2020). The uni- and multivariate models in evoTS (and paleoTS) can only be meaningfully applied to data on scale types where calculations of variances and covariances are valid. The two count traits in the stickleback data from Bell et al. (2006) are on a ratio scale (Stevens 1946), while the scale type of the pelvic score is more difficult to define. The description of how the score was applied (Bell et al. 2006, p. 567) suggests a non-linear relationship between increments of the score, which disqualifies it as a measurement on a ratio and interval scale (Stevens 1946), thus making calculations of variances and covariances non-sensical. The pelvic score was accordingly not included in the analysis. The two remaining traits were log-transformed before fitting the models. I removed one and two samples from the number of spines and number of touching pteryigiophores respectively to make samples overlap in time. The total length of the multivariate data set is 54 samples (Fig. 4).

Figure 4. Multivariate evolution in a stickleback lineage.

The vertical lines represent one standard error of the trait mean.

Fitted Models

I fitted multivariate unbiased random walks assuming either uncorrelated (only diagonal elements in R) or correlated stochastic changes (completely parameterized R) in the two traits. I also fitted different implementations of the multivariate OU model to investigate (1) if the two traits evolved independently (only diagonal elements in the A and R matrix), (2) if the traits showed evidence of independent adaptation but correlated stochastic evolution (only diagonal elements in A, but a fully parameterized R), if (3) one trait affected the optimum of the other trait (upper and lower triangular A, respectively, with a diagonal R), and, if (4) both traits affect the optimum of the other trait (fully parameterized A and only diagonal elements in R). The two last model types (3 and 4) investigate Granger causality between the two traits, i.e., whether we find evidence of changes in the traits affecting the optimum of the other trait. Complex multivariate models may have multi-peaked log-likelihood surfaces. The numerical optimization procedure was therefore initiated from 100 different starting points in parameter space for each model to minimize the risk of converging on a local peak.

Results

A multivariate OU model with independent adaptation and correlated stochastic changes showed the best relative model fit, but a model where the traits evolve independently have a similar fit according to AICc (Table 3). Models involving Granger causality (A matrices with non-zero off-diagonal elements) and the multivariate unbiased random walks are not supported. The point estimate of the half-life for the log number of spines is 1338 years according to the best model (769 generations given a generation time of two years as assumed in Hunt et al. 2008, Table 4), while the corresponding estimate for log number of touching pteryigiophores is 1276 years (638 generations) These rates of adaptation are similar to the rates reported in Hunt et al. (2008). The strength of the correlation in the stochastic trait changes is 0.48, which suggests that there may be a common, but unknown, underlying driving force for certain parts of the stochastic trait dynamics. The source of this stochasticity is not known, but genetic drift may be a contributing factor, as suggested by the univariate analyses of Hunt et al. (2008). Moreover, the presence of correlated changes in the multivariate trait dynamics aligns with research on living sticklebacks, which has found evidence of a shared genetic basis for some armor traits (e.g., Cresko et al. 2004; Leinonen et al. 2011).

Table 3.

Model fit to the multivariate stickleback sequence data. The log-likelihood (log-lik.) and the relative model fit for the candidate models fitted to the evolutionary sequence of stickleback. URW = unbiased random walk.

Model type	Multivariate model	K	Log-lik.	AICc	delta AICc	AICc weight
URW	Independent URW	4	147.479	-286.142	27.317	0.000
	Correlated evolution	5	155.559	-299.868	13.590	0.001
OU	Undependent evolution	8	165.731	-312.261	1.197	0.285
	Independent adaptation	9	167.780	-313.469	0.000	0.518
	Spines affecting touching pteryigiophores	9	166.868	-311.646	3.999	0.070
	Touching pteryigiophores affecting spines	9	166.944	-311.797	3.606	0.085
	Reciprocal effects in optima	10	167.106	-309.095	5.612	0.031

Table 4.

Maximum likelihood parameter estimates for the candidate models fitted to the multivariate evolutionary sequence of stickleback armor trait evolution. See equations and main text for definitions of the different model parameters. The numbers in parentheses are standard errors calculated from the square-root of the inverse of the diagonal of the Hessian matrix.

Multivariate model	Parameter estimates (standard error)
Independent URW	Z0.1 = 1.35 (0.02), Z0.2 = 1.39 (0.03), R1.1 = 0.09 (0.04), R2.2 = 0.14 (0.06)
Correlated evolution	Z0.1 = 1.36 (0.02), Z0.2 = 1.39 (0.03), R1.1 = 0.09 (0.00), R2.2 = 0.11 (0.00), R1.2 = 0.08 (0.00)
Independent evolution	Z0.1 = 1.37 (0.02), Z0.2 = 1.43 (0.03), R1.1 = 0.05 (0.00), R2.2 = 0.05 (0.01), A1.1 = 6.34 (1.63), A2.2 = 7.98 (1.87), θ1 = 0.80 (0.04), θ2 = 0.80 (0.04)
Independent adaptation	Z0.1 = 1.37 (0.02), Z0.2 = 1.42 (0.02), R1.1 = 0.04 (0.00), R2.2 = 0.08 (0.00), R1.2 = -0.03 (0.00), A1.1 = 6.72 (1.70), A2.2 = 8.13 (2.23), θ1 = 0.81 (0.04), θ2= 0.81 (0.04)
Spines affecting touching pteryigiophores	Z0.1 = 1.36 (0.02), Z0.2 = 1.43 (0.03), R1.1 = 0.05 (NA), R2.2 = 0.05 (NA), A1.1 = 17.70 (7.65), A2.2 = 8.84 (2.18), A1.2 = -12.72 (8.42), θ1 = 0.81 (0.03), θ2= 0.82 (0.04)
Touching pteryigiophores affecting spines	Z0.1 = 1.37 (0.02), Z0.2 = 1.43 (0.03), R1.1 = 0.05 (0.00), R2.2 = 0.07 (0.01), A1.1 = 5.44 (1.54), A2.2 = 16.61 (7.67), A2.1 = -9.78 (8.07), θ1 = 0.79 (0.05), θ2 = 0.79 (0.04)
Reciprocal effects in optima	Z0.1 = 1.37 (0.02), Z0.2 = 1.43 (0.02), R1.1 = 0.05 (-), R2.2 = 0.06 (0.01), A1.1 = 5.25 (NA), A2.2 = 15.77 (4.79), A1.2 = 1.22 (NA), A2.1 = -9.84 (NA), θ1 = 0.81 (0.04), θ2 = 0.80 (0.03)

Discussion

Analysis of evolutionary time-series in the rock record represents a unique contribution from paleontology to further the development of evolutionary biology. For example, the adaptive landscape has for decades been suggested as a conceptual bridge to bring our understanding of microevolution and macroevolution closer together (e.g., Simpson 1944; Arnold et al. 2001; Hansen 2012). Shifts in the adaptive landscape along branches of a phylogeny using Ornstein-Uhlenbeck processes are commonly explored (e.g., Ingram and Mahler 2013; Uyeda and Harmon 2014; Khabbazian et al. 2016), but whether detected shifts in optima represent sudden and major changes of the adaptive landscape or if they instead reflect cumulative changes in the position of adaptive peaks across time is poorly known (e.g., Uyeda and Harmon 2014). Inferring the dynamics of the adaptive landscape through analysis of evolutionary time-series may shed light on this and other questions regarding the dynamical nature of the adaptive landscape at different temporal scales. This potential is exemplified by the reanalysis of size evolution in the radiolarian lineage E. calvertense. The best model fitting the E. calvertense data suggests that size evolution in the lineage was affected by a major and sudden change in the adaptive landscape, while a model where the adaptive landscape changed more gradually had a poorer, but still similar, fit.

Multivariate models allow tests of a range of different hypotheses that are difficult to assess by relying on univariate models only. Two traits that appear to evolve together (i.e., show correlated evolution) over time may, for example be influenced by the same selective agent (e.g. temperature). Another possibility is that only one of the traits is directly affected by the selective agent and that changes in this trait lead to changes in the optimal value of the second trait, resulting in a somewhat lagged evolutionary response in the second trait relative to the first trait. Investigating competing explanations for multivariate trait dynamics requires a multivariate approach. The reanalysis of the stickleback data (Bell et al. 2006; Hunt et al. 2008) examined various hypotheses regarding evolution and adaptation in two armor traits. The results support independent evolution of each armor trait towards separate optima on the adaptive landscape, but a substantial portion of the stochastic changes in the two traits were correlated. The field of quantitative genetics has convincingly demonstrated how genetic correlations may affect phenotypic evolution (Lande 1979, Lande and Arnold 1983; Walsh and Blows 2009) and some armor traits in extant threespine stickleback have been found to covary genetically (e.g., Cresko et al. 2004; Leinonen et al. 2011), suggesting these traits are likely to have a reduced capacity to evolve completely independent of each other. One possible interpretation of the detected non-independence in part of the evolution of the two armor traits is therefore due to a shared genetic background. An alternative and not mutually exclusive explanation is that these two armor traits have experienced correlated selection.

It is important to note that the interpretation of the various univariate and multivariate trait models in evoTS (and paleoTS) may vary depending on the time interval covered by the analyzed data. For example, the OU process has been used to describe microevolutionary changes in a population close to a fixed peak in the adaptive landscape (Lande 1976; Hansen and Martins 1996), but it is also commonly used to model evolution within and between adaptive zones on among-species comparative data (e.g. Mahler et al. 2013; Moen et al. 2016; Toljagic et al. 2018). Fitting OU models to evolutionary time-series of modern lineages where the data have a generational resolution allows for the estimation of microevolutionary parameters and the development of a process-based interpretation of the trait dynamics based on the parameters in the OU model (see Lo Cascio Sætre et al. 2017 for an example). The fossil record may not always have the necessary resolution for interpreting the model directly in terms of microevolutionary processes, in which case a more phenomenological interpretation of the fitted models may be more appropriate. For example, interpreting the stochastic part of the trait dynamics in an OU model as primarily a result of genetic drift may be more suitable when fitting the model to modern data with high time resolution than fitting it to fossil data with lower time resolution. However, there is no strict boundary between when a process-based and when a more phenomenological interpretation is most suitable. In their analysis of the same stickleback armor traits analyzed in this study, Hunt et al. (2008) demonstrated that microevolutionary parameters can be meaningfully estimated from the OU model. Therefore, the best way to interpret the model parameters should be assessed on a case-by-case basis.

The number of models and available software for conducting phylogenetic comparative analyses have steadily increased for more than 30 years, giving ample opportunities for exploring a large range of hypotheses of trait dynamics on macroevolutionary time scales (see e.g., Pennell and Harmon 2013). Analysis of evolutionary time-series has not experienced a similar momentum, likely due to the smaller number of available evolutionary time-series relative to phylogenetic comparative data. The R package paleoTS has for a long time been the most popular software for fitting models to evolutionary time-series. The implemented univariate and multivariate models in evoTS extends the model options available in paleoTS. However, evoTS is not the only software that allows fitting multivariate models to evolutionary time-series. layeranalyzer (Reitan and Liow 2019) is a tool that can be used to explore correlations and causal relationships among variables in time-series and can fit most of the models implemented in evoTS. mvmorph (Clavel et al. 2015) is mainly targeted towards analysis of multivariate phylogenetic comparative data, but several of the implemented models can also be fitted to time-series. The main advantage of evoTS is that it works as an extension of the much-used paleoTS framework and is therefore specifically tailored towards analysis of evolutionary time-series. The combined suite of univariate and multivariate models in paleoTS and evoTS is also not found in any alternative software.

Connecting known microevolutionary processes to macroevolutionary patterns remain a central challenge in biology (e.g., Arnold et al. 2001; Jablonski 2000; 2017; Hansen 2012). While data on generational changes and among-taxa differences are readily available for many organismal groups, information on how lineages evolve in-between micro- and macroevolutionary time scales is rare in comparison. Furthermore, how to interpret evolutionary change recorded in the rock record has been debated for decades (e.g., Eldredge and Gould 1972; Charlesworth et al. 1982; Gingerich 1984; 2019; Stanley 1985; Bookstein 1987; Hunt 2007; Voje 2016), and the nature of evolutionary change within lineages remains controversial (Lieberman and Eldredge 2014; Pennell et al. 2014a, 2014b; Venditti and Pagel 2014). Investigating and comparing a larger range of models and hypotheses when analyzing phenotypic change within lineages may contribute to these ongoing discussions.

Data Availability Statement

Data and R code available from the Dryad Digital Repository: https://doi.org/10.5061/dryad.c59zw3rcp.