This vignette provides an introduction to the stan_jm
modelling function in the rstanarm package. The stan_jm
function allows the user to estimate a shared parameter joint model for longitudinal and timetoevent data under a Bayesian framework.
Joint modelling can be broadly defined as the simultaneous estimation of two or more statistical models which traditionally would have been separately estimated. When we refer to a shared parameter joint model for longitudinal and timetoevent data, we generally mean the joint estimation of: 1) a longitudinal mixed effects model which analyses patterns of change in an outcome variable that has been measured repeatedly over time (for example, a clinical biomarker) and 2) a survival or timetoevent model which analyses the time until an event of interest occurs (for example, death or disease progression). Joint estimation of these socalled “submodels” is achieved by assuming they are correlated via individualspecific parameters (i.e. individuallevel random effects).
Over the last two decades the joint modelling of longitudinal and timetoevent data has received a significant amount of attention [15]. Methodological developments in the area have been motivated by a growing awareness of the benefits that a joint modelling approach can provide. In clinical or epidemiological research it is common for a clinical biomarker to be repeatedly measured over time on a given patient. In addition, it is common for timetoevent data, such as the patientspecific time from a defined origin (e.g. time of diagnosis of a disease) until a terminating clinical event such as death or disease progression to also be collected. The figure below shows observed longitudinal measurements (i.e. observed “trajectories”) of log serum bilirubin for a small sample of patients with primary biliary cirrhosis. Solid lines are used for those patients who were still alive at the end of follow up, while dashed lines are used for those patients who died. From the plots, we can observe betweenpatient variation in the longitudinal trajectories for log serum bilirubin, with some patients showing an increase in the biomarker over time, others decreasing, and some remaining stable. Moreover, there is variation between patients in terms of the frequency and timing of the longitudinal measurements.
From the perspective of clinical risk prediction, we may be interested in asking whether the betweenpatient variation in the log serum bilirubin trajectories provides meaningful prognostic information that can help us differentiate patients with regard to some clinical event of interest, such as death. Alternatively, from an epidemiological perspective we may wish to explore the potential for etiological associations between changes in log serum bilirubin and mortality. Joint modelling approaches provide us with a framework under which we can begin to answer these types of clinical and epidemiological questions.
More formally, the motivations for undertaking a joint modelling analysis of longitudinal and timetoevent data might include one or more of the following:
One may be interested in how underlying changes in the biomarker influence the occurrence of the event. However, including the observed biomarker measurements directly into a timetoevent model as timevarying covariates poses several problems. For example, if the widely used Cox proportional hazards model is assumed for the timetoevent model then biomarker measurements need to be available for all patients at all failure times, which is unlikely to be the case [3]. If simple methods of imputation are used, such as the “last observation carried forward” method, then these are likely to induce bias [6]. Furthermore, the observed biomarker measurements may be subject to measurement error and therefore their inclusion as timevarying covariates may result in biased and inefficient estimates. In most cases, the measurement error will result in parameter estimates which are shrunk towards the null [7]. On the other hand, joint modelling approaches allow us to estimate the association between the biomarker (or some function of the biomarker trajectory, such as rate of change in the biomarker) and the risk of the event, whilst allowing for both the discrete time and measurementerror aspects of the observed biomarker.
One may be interested primarily in the evolution of the clinical biomarker but may wish to account for what is known as informative dropout. If the value of future (unobserved) biomarker measurements are related to the occurrence of the terminating event, then those unobserved biomarker measurements will be “missing not at random” [8,9]. In other words, biomarker measurements for patients who have an event will differ from those who do not have an event. Under these circumstances, inference based solely on observed measurements of the biomarker will be subject to bias. A joint modelling approach can help to adjust for informative dropout and has been shown to reduce bias in the estimated parameters associated with longitudinal changes in the biomarker [1,9,10].
Joint models are naturally suited to the task of dynamic risk prediction. For example, joint modelling approaches have been used to develop prognostic models where predictions of event risk can be updated as new longitudinal biomarker measurements become available. Taylor et al. [11] jointly modelled longitudinal measurements of the prostate specific antigen (PSA) and time to clinical recurrence of prostate cancer. The joint model was then used to develop a webbased calculator which could provide realtime predictions of the probability of recurrence based on a patient’s up to date PSA measurements.
In this vignette, we describe the rstanarm package’s stan_jm
modelling function. This modelling function allows users to fit a shared parameter joint model for longitudinal and timetoevent data under a Bayesian framework, with the backend estimation carried out using Stan. In Section 2 we describe the formulation of the joint model used by stan_jm
. In Section 3 we present a variety of examples showing the usage of stan_jm
.
Note that some aspects of the estimation are covered in other vignettes, such as the priors vignette which contains details on the prior distributions available for regression coefficients.
A shared parameter joint model consists of related submodels which are specified separately for each of the longitudinal and timetoevent outcomes. These are therefore commonly referred to as the longitudinal submodel(s) and the event submodel. The longitudinal and event submodels are linked using shared individualspecific parameters, which can be parameterised in a number of ways. We describe each of these submodels below.
We assume \(y_{ijm}(t) = y_{im}(t_{ij})\) corresponds to the observed value of the \(m^{th}\) \((m = 1,...,M)\) biomarker for individual \(i\) \((i = 1,...,N)\) taken at time point \(t_{ij}\), \(j = 1,...,n_{im}\). We specify a (multivariate) generalised linear mixed model that assumes \(y_{ijm}(t)\) follows a distribution in the exponential family with mean \(\mu_{ijm}(t)\) and linear predictor
\[ \eta_{ijm}(t) = g_m(\mu_{ijm}(t)) = \boldsymbol{x}^T_{ijm}(t) \boldsymbol{\beta}_m + \boldsymbol{z}^T_{ijm}(t) \boldsymbol{b}_{im} \]
where \(\boldsymbol{x}^T_{ijm}(t)\) and \(\boldsymbol{z}^T_{ijm}(t)\) are both rowvectors of covariates (which likely include some function of time, for example a linear slope, cubic splines, or polynomial terms) with associated vectors of fixed and individualspecific parameters \(\boldsymbol{\beta}_m\) and \(\boldsymbol{b}_{im}\), respectively, and \(g_m\) is some known link function. The distribution and link function are allowed to differ over the \(M\) longitudinal submodels. We let the vector \(\boldsymbol{\beta} = \{ \boldsymbol{\beta}_m ; m = 1,...,M\}\) denote the collection of populationlevel parameters across the \(M\) longitudinal submodels. We assume that the dependence across the different longitudinal submodels (i.e. the correlation between the different longitudinal biomarkers) is captured through a shared multivariate normal distribution for the individualspecific parameters; that is, we assume
\[ \begin{pmatrix} \boldsymbol{b}_{i1} \\ \vdots \\ \boldsymbol{b}_{iM} \end{pmatrix} = \boldsymbol{b}_i \sim \mathsf{Normal} \left( 0 , \boldsymbol{\Sigma} \right) \]
for some unstructured variancecovariance matrix \(\boldsymbol{\Sigma}\).
We assume that we also observe an event time \(T_i = \mathsf{min} \left( T^*_i , C_i \right)\) where \(T^*_i\) denotes the socalled “true” event time for individual \(i\) (potentially unobserved) and \(C_i\) denotes the censoring time. We define an event indicator \(d_i = I(T^*_i \leq C_i)\). We then model the hazard of the event using a parametric proportional hazards regression model of the form
\[ h_i(t) = h_0(t; \boldsymbol{\omega}) \mathsf{exp} \left( \boldsymbol{w}^T_i(t) \boldsymbol{\gamma} + \sum_{m=1}^M \sum_{q=1}^{Q_m} f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) \right) \]
where \(h_i(t)\) is the hazard of the event for individual \(i\) at time \(t\), \(h_0(t; \boldsymbol{\omega})\) is the baseline hazard at time \(t\) given parameters \(\boldsymbol{\omega}\), \(\boldsymbol{w}^T_i(t)\) is a rowvector of individualspecific covariates (possibly timedependent) with an associated vector of regression coefficients \(\boldsymbol{\gamma}\) (log hazard ratios), \(f_{mq}(.)\) are a set of known functions for \(m=1,...,M\) and \(q=1,...,Q_m\), and the \(\alpha_{mq}\) are regression coefficients (log hazard ratios).
The longitudinal and event submodels are assumed to be related via an “association structure”, which is a set of functions each \(\{ f_{mq} ; m = 1,...,M, q = 1,...,Q_m \}\) that may each be conditional on the populationlevel parameters from the longitudinal submodel \(\boldsymbol{\beta}\), the individualspecific parameters \(\boldsymbol{b}_{i}\), and the populationlevel parameters \(\alpha_{mq}\) for \(m=1,...,M\) and \(q=1,...,Q_m\). That is, the association structure of the joint model is captured via the \(\sum_{m=1}^M \sum_{q=1}^{Q_m} f_{mq}(\boldsymbol{\beta}_m, \boldsymbol{b}_{im}, \alpha_{mq}; t)\) term in the linear predictor of the event submodel. The \(\alpha_{mq}\) are referred to as the “association parameters” since they quantify the strength of the association between the longitudinal and event processes. The various ways in which the association structure can be are described in the next section.
The probability of individual \(i\) still being eventfree at time \(t\), often referred to as the “survival probability”, is defined as
\[ S_i(t) = \text{Prob} \Big( T_i^* \geq t \Big) = \exp \Big( H_i(t) \Big) \]
where \(H_i(t) = \int_{s=0}^t h_i(s) ds\) is the cumulative hazard for individual \(i\).
We assume that the baseline hazard \(h_0(t; \boldsymbol{\omega})\) is modelled parametrically. In the stan_jm
modelling function the baseline hazard be specified as either: an approximation using Bsplines on the log hazard scale (the default); a Weibull distribution; or an approximation using a piecewise constant function on the log hazard scale (sometimes referred to as piecewise exponential). The choice of baseline hazard can be made via the basehaz
argument. In the case of the Bsplines or piecewise constant baseline hazard, the user can control the flexibility by specifying the knots or degrees of freedom via the basehaz_ops
argument. (Note that currently there is slightly limited postestimation functionality available for models estimated with a piecewise constant baseline hazard, so this is perhaps the least preferable choice).
As mentioned in the previous section, the dependence between the longitudinal and event submodels is captured through the association structure, which can be specified in a number of ways. The simplest association structure is likely to be
\[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{im}, \alpha_{mq}; t) = \alpha_{mq} \eta_{im}(t) \]
and this is often referred to as a current value association structure since it assumes that the log hazard of the event at time \(t\) is linearly associated with the value of the longitudinal submodel’s linear predictor also evaluated at time \(t\). This is the most common association structure used in the joint modelling literature to date. In the situation where the longitudinal submodel is based on an identity link function and normal error distribution (i.e. a linear mixed model) the current value association structure can be viewed as a method for including the underlying “true” value of the biomarker as a timevarying covariate in the event submodel.^{1}
However, other association structures are also possible. For example, we could assume the log hazard of the event is linearly associated with the current slope (i.e. rate of change) of the longitudinal submodel’s linear predictor, that is
\[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \frac{d\eta_{im}(t)}{dt} \]
There are in fact a whole range of possible association structures, many of which have been discussed in the literature [1416].
The stan_jm
modelling function in the rstanarm package allows for the following association structures, which are specified via the assoc
argument:
Current value (of the linear predictor or expected value) \[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \eta_{im}(t) \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \mu_{im}(t) \]
Current slope (of the linear predictor or expected value) \[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \frac{d\eta_{im}(t)}{dt} \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \frac{d\mu_{im}(t)}{dt} \]
Area under the curve (of the linear predictor or expected value) \[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \int_0^t \eta_{im}(u) du \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \int_0^t \mu_{im}(u) du \]
Interactions between different biomarkers \[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \eta_{im}(t) \eta_{im'}(t) \text{ for some } m = m' \text{ or } m \neq m' \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \eta_{im}(t) \mu_{im'}(t) \text{ for some } m = m' \text{ or } m \neq m' \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \mu_{im}(t) \mu_{im'}(t) \text{ for some } m = m' \text{ or } m \neq m' \]
Interactions between the biomarker (or it’s slope) and observed data \[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} c_{i}(t) \eta_{im}(t) \text{ for some covariate value } c_{i}(t) \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} c_{i}(t) \mu_{im}(t) \text{ for some covariate value } c_{i}(t) \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} c_{i}(t) \frac{d\eta_{im}(t)}{dt} \text{ for some covariate value } c_{i}(t) \\ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} c_{i}(t) \frac{d\mu_{im}(t)}{dt} \text{ for some covariate value } c_{i}(t) \]
As well as using lagged values for any of the above. That is, replacing \(t\) with \(tu\) where \(u\) is some lag time, such that the hazard of the event at time \(t\) is assumed to be associated with some function of the longitudinal submodel parameters at time \(tu\).
Lastly, we can specify some timeindependent function of the random effects, possibly including the fixed effect component. For example,
\[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \boldsymbol{b}_{im0} \]
or
\[ f_{mq}(\boldsymbol{\beta}, \boldsymbol{b}_{i}, \alpha_{mq}; t) = \alpha_{mq} \Big( \boldsymbol{\beta}_{m0} + \boldsymbol{b}_{im0} \Big) \]
where \(\boldsymbol{\beta}_{m0}\) is the populationlevel intercept for the \(m^{th}\) longitudinal submodel and \(\boldsymbol{b}_{im0}\) is the \(i^{th}\) individual’s random deviation from the populationlevel intercept for the \(m^{th}\) longitudinal submodel.
Note that more than one association structure can be specified, however, not all possible combinations are allowed. Moreover, if you are fitting a multivariate joint model (i.e. more than one longitudinal outcome) then you can optionally choose to use a different association structure(s) for linking each longitudinal submodel to the event submodel. To do this you can pass a list of length \(M\) to the assoc
argument.
Here we define a set of assumptions for the multivariate shared parameter joint model.
The socalled conditional independence assumption of the shared parameter joint model postulates
\[ y_{im}(t) \perp y_{im'}(t) \mid \boldsymbol{b}_i, \boldsymbol{\theta} \\ y_{im}(t) \perp y_{im}(t') \mid \boldsymbol{b}_i, \boldsymbol{\theta} \\ y_{im}(t) \perp T_i^* \mid \boldsymbol{b}_i, \boldsymbol{\theta} \]
for some \(m \neq m'\) and \(t \neq t'\), and where \(\boldsymbol{\theta}\) denotes the combined vector of all remaining populationlevel parameters in the model. That is, conditional on the individualspecific parameters \(\boldsymbol{b}_i\) and populationlevel parameters \(\boldsymbol{\theta}\), the following are assumed: (i) any biomarker measurement for individual \(i\) is independent of that individual’s true event time \(T_i^*\); (ii) any two measurements of the \(m^{th}\) biomarker taken on the \(i^{th}\) individual at two distinct time points \(t\) and \(t'\) (i.e. longitudinal or repeated measurements) are independent of one another; and (iii) any two measurements of two different biomarkers, taken on the \(i^{th}\) individual at some time point \(t\) are independent of one another. These conditional independence assumptions allow for a convenient factorisation of the full likelihood for joint model into the likelihoods for each of the component parts (i.e. the likelihood for the longitudinal submodel, the likelihood for the event submodel, and the likelihood for the distribution of the individualspecific parameters), which facilitates the estimation of the model.
Moreover, we require two additional assumptions: (i) that the censoring process for the event outcome is independent of the true event time, that is \(C_i \perp T_i^* \mid \boldsymbol{\theta}\) (i.e. uninformative censoring); and (ii) that the visiting process by which the observation times \(t_{ijm}\) are determined is independent of the true event time \(T_i^*\) and all missing future unobserved longitudinal biomarker measurements.
Under the conditional independence assumption, the log posterior for the \(i^{th}\) individual can be specified as
\[ \log p(\boldsymbol{\theta}, \boldsymbol{b}_{i} \mid \boldsymbol{y}_{i}, T_i, d_i) \propto \log \Bigg[ \Bigg( \prod_{m=1}^M \prod_{j=1}^{n_i} p(y_{ijm}(t) \mid \boldsymbol{b}_{i}, \boldsymbol{\theta}) \Bigg) p(T_i, d_i \mid \boldsymbol{b}_{i}, \boldsymbol{\theta}) p(\boldsymbol{b}_{i} \mid \boldsymbol{\theta}) p(\boldsymbol{\theta}) \Bigg] \]
where \(\boldsymbol{y}_i = \{ y_{ijm}(t); j = 1,...,n_i, m = 1,...,M \}\) denotes the collection of longitudinal biomarker data for individual \(i\) and \(\boldsymbol{\theta}\) denotes all remaining populationlevel parameters in the model.
We can rewrite this log posterior as
\[ \log p(\boldsymbol{\theta}, \boldsymbol{b}_{i} \mid \boldsymbol{y}_{i}, T_i, d_i) \propto \Bigg( \sum_{m=1}^M \sum_{j=1}^{n_i} \log p(y_{ijm}(t) \mid \boldsymbol{b}_{i}, \boldsymbol{\theta}) \Bigg) + \log p(T_i, d_i \mid \boldsymbol{b}_{i}, \boldsymbol{\theta}) + \log p(\boldsymbol{b}_{i} \mid \boldsymbol{\theta}) + \log p(\boldsymbol{\theta}) \]
where \(\sum_{j=1}^{n_{im}} \log p(y_{ijm} \mid \boldsymbol{b}_{i}, \boldsymbol{\theta})\) is the log likelihood for the \(m^{th}\) longitudinal submodel, \(\log p(T_i, d_i \mid \boldsymbol{b}_{i}, \boldsymbol{\theta})\) is the log likelihood for the event submodel, \(\log p(\boldsymbol{b}_{i} \mid \boldsymbol{\theta})\) is the log likelihood for the distribution of the groupspecific parameters (i.e. random effects), and \(\log p(\boldsymbol{\theta})\) represents the log likelihood for the joint prior distribution across all remaining unknown parameters.^{2}
We can rewrite the log likelihood for the event submodel as
\[ \log p(T_i, d_i \mid \boldsymbol{b}_{i}, \boldsymbol{\theta}) = d_i * \log h_i(T_i)  \int_0^{T_i} h_i(s) ds \]
and then use GaussKronrod quadrature with \(Q\) nodes to approximate \(\int_0^{T_i} h_i(s) ds\), such that
\[ \int_0^{T_i} h_i(s) ds \approx \frac{T_i}{2} \sum_{q=1}^{Q} w_q h_i \bigg( \frac{T_i(1+s_q)}{2} \bigg) \]
where \(w_q\) and \(s_q\), respectively, are the standardised weights and locations (“abscissa”) for quadrature node \(q\) \((q=1,...,Q)\) [17]. The default for the stan_jm
modelling function is to use \(Q=15\) quadrature nodes, however if the user wishes, they can choose between \(Q=15\), \(11\), or \(7\) quadrature nodes (specified via the qnodes
argument).
Therefore, once we have an individual’s event time \(T_i\) we can evaluate the design matrices for the event submodel and longitudinal submodels at the \(Q+1\) necessary time points (which are the event time \(T_i\) and the quadrature points \(\frac{T_i(1+s_q)}{2}\) for \(q=1,...,Q\)) and then pass these to Stan’s data block. We can then evaluate the log likelihood for the event submodel by simply calculating the hazard \(h_i(t)\) at those \(Q+1\) time points and summing the quantities appropriately. This calculation will need to be performed each time we iterate through Stan’s model block. A simplified example of the underlying Stan code used to fit the joint model can be found in Brilleman et al. (2018) [12].
Before discussing the methods by which we can generate posterior predictions, first let us define some additional relevant quantities. Let \(\mathcal{D} = \{ \boldsymbol{y}_i, T_i, d_i; i = 1,...,N \}\) be the entire collection of outcome data in the sample. We will refer to this sample as the “training data”. Let \(T_{max} = \max \{ T_i; i = 1,...,N \}\) denote the maximum event or censoring time across the \(i = 1,...,N\) individuals in our training data.
We can generate posterior predictions for the longitudinal and timetoevent outcomes in the following manner. For the \(i^{th}\) individual in our training data, a predicted value for the \(m^{th}\) longitudinal biomarker at time \(t\), denoted \(y^*_{im}(t)\), can be generated from the posterior predictive distribution
\[ p \Big( y^{*}_{im}(t) \mid \mathcal{D} \Big) = \int \int p \Big( y^{*}_{im}(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_i \Big) p \Big( \boldsymbol{\theta}, \boldsymbol{b}_i \mid \mathcal{D} \Big) \space d \boldsymbol{b}_i \space d \boldsymbol{\theta} \]
and, similarly, a predicted probability of the \(i^{th}\) individual being eventfree at time \(t\), denoted \(S^*_i(t)\), can be generated from the posterior predictive distribution
\[ p \Big( S^{*}_{i}(t) \mid \mathcal{D} \Big) = \int \int p \Big( S^{*}_i(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_i \Big) p \Big( \boldsymbol{\theta}, \boldsymbol{b}_i \mid \mathcal{D} \Big) d \boldsymbol{b}_i \space d \boldsymbol{\theta} \]
Note that for simplicity we have ignored the implicit conditioning on covariates; \(\boldsymbol{x}_{im}(t)\) and \(\boldsymbol{z}_{im}(t)\), for \(m = 1,...,M\), and \(\boldsymbol{w}_{i}(t)\). Since individual \(i\) is included in the training data, it is easy for us to approximate these posterior predictive distributions by drawing from \(p(y^{*}_{im}(t) \mid \boldsymbol{\theta}^{(l)}, \boldsymbol{b}_i^{(l)})\) and \(p(S^{*}_i(t) \mid \boldsymbol{\theta}^{(l)}, \boldsymbol{b}_i^{(l)})\) where \(\boldsymbol{\theta}^{(l)}\) and \(\boldsymbol{b}_i^{(l)}\) are the \(l^{th}\) \((l = 1,...,L)\) MCMC draws from the joint posterior distribution \(p(\boldsymbol{\theta}, \boldsymbol{b}_i \mid \mathcal{D})\).
These draws from the posterior predictive distributions can be used for assessing the fit of the model. For example,
the draws from \(p(y^{*}_{im}(t) \mid \mathcal{D})\) for \(0 \leq t \leq T_i\) can be used to evaluate the fit of the longitudinal trajectory for the \(m^{th}\) biomarker for the \(i^{th}\) individual, and
the draws from \(p(S^{*}_{i}(t) \mid \mathcal{D})\) for \(0 \leq t \leq T_{max}\) can be averaged across the \(N\) individuals to obtain a standardised survival curve (discussed in greater detail in later sections) which can then be compared to the observed survival curve, for example, the KaplanMeier curve.
However, given that we know the event or censoring time for each individual in our training data, it may make more sense to consider what will happen to censored individuals in our study when we look beyond their last known survival time (i.e. extrapolation).
For an individual \(i\), who was in our training data, and who was known to be eventfree up until their censoring time \(C_i\), we wish to draw from the conditional posterior predictive distribution for their longitudinal outcome at some time \(t > C_i\), that is
\[ p \Big( y^{*}_{im}(t) \mid \mathcal{D}, t > C_i \Big) = \int \int p \Big( y^{*}_{im}(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_i, t > C_i \Big) p \Big( \boldsymbol{\theta}, \boldsymbol{b}_i \mid \mathcal{D} \Big) d \boldsymbol{b}_i \space d \boldsymbol{\theta} \]
and the conditional posterior predictive distribution for their survival probability at some time \(t > C_i\), that is
\[ \begin{aligned} p \Big( S^{*}_{i}(t) \mid \mathcal{D}, t > C_i, T_i^* > C_i \Big) & = \frac {p \Big( S^{*}_{i}(t) \mid \mathcal{D} \Big)} {p \Big( S^{*}_{i}(C_i) \mid \mathcal{D} \Big)} \\ & = \int \int \frac {p \Big( S^{*}_i(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_i \Big)} {p \Big( S^{*}_i(C_i) \mid \boldsymbol{\theta}, \boldsymbol{b}_i \Big)} \space p \Big( \boldsymbol{\theta}, \boldsymbol{b}_i \mid \mathcal{D} \Big) d \boldsymbol{b}_i \space d \boldsymbol{\theta} \end{aligned} \]
These draws from the conditional posterior predictive distributions can be used to extrapolate into the future for individual \(i\), conditional on their longitudinal biomarker data collected between baseline and their censoring time \(C_i\). For example,
the draws from \(p(y^{*}_{im}(t) \mid \mathcal{D}, t > C_i)\) for \(C_i \leq t \leq T_{max}\) can be used to show the forecasted longitudinal trajectory for the \(m^{th}\) biomarker for the \(i^{th}\) individual, and
the draws from \(p(S^{*}_{i}(t) \mid \mathcal{D}, t > C_i, T_i^* > C_i))\) for \(C_i \leq t \leq T_{max}\) can be used to show the estimated conditional probability of individual \(i\) remaining eventfree into the future.
TBC. Describe dynamic predictions under the framework of Rizopoulos (2011) [18]. These types of individualspecific predictions can be obtained using the posterior_traj
and posterior_survfit
functions by providing prediction data and specifying dynamic = TRUE
(which is the default); see the examples provided below.
We can also generate posterior predictions for the longitudinal and timetoevent outcomes that do not require any conditioning on observed outcome data for a specific individual. Here, we will discuss two ways in which this can be done.
The first way is to “marginalise” over the distribution of the individualspecific parameters. We wish to generate a predicted value for the \(m^{th}\) longitudinal biomarker at time \(t\) for a new individual \(k\) for whom we do not have any observed data. We will denote this prediction \(y^*_{km}(t)\) and note that it can be generated from the posterior predictive distribution for the longitudinal outcome
\[ \begin{aligned} p \Big( y^{*}_{km}(t) \mid \mathcal{D} \Big) & = \int \int p \Big( y^{*}_{km}(t) \mid \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \Big) p \Big( \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \mid \mathcal{D} \Big) \space d \boldsymbol{\tilde{b}}_{k} \space d \boldsymbol{\theta} \\ & = \int \int p \Big( y^{*}_{km}(t) \mid \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \Big) p \Big( \boldsymbol{\tilde{b}}_k \mid \boldsymbol{\theta} \Big) p \Big( \boldsymbol{\theta} \mid \mathcal{D} \Big) \space d \boldsymbol{\tilde{b}}_{k} \space d \boldsymbol{\theta} \end{aligned} \]
and similarly for the survival probability
\[ \begin{aligned} p \Big( S^{*}_{k}(t) \mid \mathcal{D} \Big) & = \int \int p \Big( S^{*}_k(t) \mid \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \Big) p \Big( \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \mid \mathcal{D} \Big) d \boldsymbol{b}_k \space d \boldsymbol{\theta} \\ & = \int \int p \Big( S^{*}_k(t) \mid \boldsymbol{\theta}, \boldsymbol{\tilde{b}}_k \Big) p \Big( \boldsymbol{\tilde{b}}_k \mid \boldsymbol{\theta} \Big) p \Big( \boldsymbol{\theta} \mid \mathcal{D} \Big) d \boldsymbol{b}_k \space d \boldsymbol{\theta} \\ \end{aligned} \]
We can obtain draws for \(\boldsymbol{\tilde{b}}_k\) in the same manner as for the individualspecific parameters \(\boldsymbol{b}_i\). That is, at the \(l^{th}\) iteration of the MCMC sampler we draw \(\boldsymbol{\tilde{b}}_k^{(l)}\) and store it^{3}. However, individual \(k\) did not provide any contribution to the training data and so we are effectively taking random draws from the posterior distribution for the individualspecific parameters. We are therefore effectively marginalising over the distribution of the groupspecific coefficients when we obtain predictions using the draws \(\boldsymbol{\tilde{b}}_k^{(l)}\) fro \(l = 1,\dots,L\). In other words, we are predicting for a new individual whom we have no information except that they are drawn from the same population as the \(i = 1,...,N\) individuals in the training data. Because these predictions will incorporate all the uncertainty associated with betweenindividual variation our 95% credible intervals are likely to be very wide. These types of marginal predictions can be obtained using the posterior_traj
and posterior_survfit
functions by providing prediction data and specifying dynamic = FALSE
; see the examples provided below.
The second way is to effectively ignore the grouplevel structure in the model. That is, to only predict with only the populationlevel parameters contributing to the model. For example, under a identity link function and normal error distribution (i.e. linear mixed effect longitudinal submodel), we would obtain draws from the distribution \(y^{(l)}_{km}(t) \sim N \Big( \boldsymbol{x}^T_{km}(t) \boldsymbol{\beta}_m^{(l)}, \sigma_m^{(l)} \Big)\) where \(\boldsymbol{\beta}_m^{(l)}\) and \(\sigma_m^{(l)}\) are the populationlevel parameters and residual error standard deviation, respectively, for the \(l^{th}\) draw of the MCMC samples. However, referring to this as a “marginal” prediction is somewhat misleading since we are not explicitly conditioning on the individualspecific parameters but we are implicitly assuming that we know they are equal to zero with absolute certainty. That is, we are actually drawing from the posterior predictive distribution for the longitudinal outcome
\[ \begin{aligned} p \Big( y^{*}_{km}(t) \mid \mathcal{D} \Big) & = \int p \Big( y^{*}_{km}(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_k = 0 \Big) p \Big( \boldsymbol{\theta} \mid \mathcal{D} \Big) d \boldsymbol{\theta} \\ \end{aligned} \] and similarly for the survival probability
\[ p \Big( S^{*}_{k}(t) \mid \mathcal{D} \Big) = \int p \Big( S^{*}_k(t) \mid \boldsymbol{\theta}, \boldsymbol{b}_k = 0 \Big) p \Big( \boldsymbol{\theta} \mid \mathcal{D} \Big) d \boldsymbol{\theta} \\ \]
These types of socalled “marginal” predictions can not currently be obtained using the posterior_traj
and posterior_survfit
functions.
All of the previously discussed populationlevel (i.e. marginal) predictions assumed implicit conditioning on some covariate values for the longitudinal submodel, \(\boldsymbol{x}_{im}(t)\) and \(\boldsymbol{z}_{im}(t)\) for \(m = 1,...,M\), and for the event submodel, \(\boldsymbol{w}_{i}(t)\). Even though we marginalise over the distribution of the individualspecific parameters we were still assuming that we obtained predictions for some known values of the covariates. However, sometimes we wish to marginalise (i.e. average) over the observed distribution of covariates as well. Here we discuss a method by which we can do that for the predicted survival probabilities.
At any time \(t\), it is possible to obtain a standardised survival probability by averaging the individualspecific survival probabilities. That is, we can obtain
\[ S^*(t) = \frac{\sum_{i=1}^{N^{pred}} S_i^*(t)}{N^{pred}} \]
where \(S_i^*(t)\) is the predicted survival probability for individual \(i\) (\(i = 1,\dots,N^{pred}\) at time \(t\), and \(N^{pred}\) is the number of individuals included in the prediction dataset. We refer to these predictions as standardised survival probabilities.
Note however, that if \(N_{pred}\) is not sufficiently large (e.g. we pass new data with just 2 individuals, say) then marginalising over their covariate distribution may not be meaningful and, similarly, their joint random effects distribution may be a poor representation of the random effects distribution for the entire population. It is better to calculate these standardised survival probabilities using where, say, \(N^{pred}\) is equal to the total number of individuals in the training data.
TBC.
TBC.
TBC.
We use the Mayo Clinic’s primary biliary cirrhosis (PBC) dataset in the examples below. The dataset contains 312 individuals with primary biliary cirrhosis who participated in a randomised placebo controlled trial of Dpenicillamine conducted at the Mayo Clinic between 1974 and 1984 [19]. However, to ensure the examples run quickly, we use a small random subset of just 40 patients from the full data.
These example data are contained in two separate data frames. The first data frame contains multiplerow per patient longitudinal biomarker information, as shown in
id age sex trt year logBili albumin platelet
1 1 58.76523 f 1 0.0000000 2.67414865 2.60 190
2 1 58.76523 f 1 0.5256674 3.05870707 2.94 183
3 2 56.44627 f 1 0.0000000 0.09531018 4.14 221
4 2 56.44627 f 1 0.4982888 0.22314355 3.60 188
5 2 56.44627 f 1 0.9993155 0.00000000 3.55 161
6 2 56.44627 f 1 2.1026694 0.64185389 3.92 122
while the second data frame contains singlerow per patient survival information, as shown in
id age sex trt futimeYears status death
1 1 58.76523 f 1 1.095140 2 1
3 2 56.44627 f 1 14.151951 0 0
12 3 70.07255 m 1 2.770705 2 1
16 4 54.74059 f 1 5.270363 2 1
23 5 38.10541 f 0 4.120465 1 0
29 6 66.25873 f 0 6.852841 2 1
The variables included across the two datasets can be defined as follows:
age
in yearsalbumin
serum albumin (g/dl)logBili
logarithm of serum bilirubindeath
indicator of death at endpointfutimeYears
time (in years) between baseline and the earliest of death, transplantion or censoringid
numeric ID unique to each individualplatelet
platelet countsex
gender (m = male, f = female)status
status at endpoint (0 = censored, 1 = transplant, 2 = dead)trt
binary treatment code (0 = placebo, 1 = Dpenicillamine)year
time (in years) of the longitudinal measurements, taken as time since baseline)A description of the example datasets can be found by accessing the following help documentation:
In this example we fit a simple univariate joint model, with one normally distributed longitudinal marker, an association structure based on the current value of the linear predictor, and Bsplines baseline hazard. To fit the model we use the joint (longitudinal and timetoevent) modelling function in the rstanarm package: stan_jm
. When calling stan_jm
we must, at a minimum, specify a formula object for each of the longitudinal and event submodels (through the arguments formulaLong
and formulaEvent
), the data frames which contain the variables for each of the the longitudinal and event submodels (through the arguments dataLong
and dataEvent
), and the name of the variable representing time in the longitudinal submodel (through the argument time_var
).
The formula for the longitudinal submodel is specified using the lme4 package formula style. That is y ~ x + (random_effects  grouping_factor)
. In this example we specify that log serum bilirubin (logBili
) follows a subjectspecific linear trajectory. To do this we include a fixed intercept and fixed slope (year
), as well as a random intercept and random slope for each subject id
((year  id)
).
The formula for the event submodel is specified using the survival package formula style. That is, the outcome of the left of the ~
needs to be of the format Surv(event_time, event_indicator)
for single row per individual data, or Surv(start_time, stop_time, event_indicator)
for multiple row per individual data. The latter allows for exogenous timevarying covariates to be included in the event submodel. In this example we assume that the log hazard of death is linearly related to gender (sex
) and an indicator of treatment with Dpenicillamine (trt
).
library(rstanarm)
mod1 < stan_jm(formulaLong = logBili ~ sex + trt + year + (year  id),
dataLong = pbcLong,
formulaEvent = survival::Surv(futimeYears, death) ~ sex + trt,
dataEvent = pbcSurv,
time_var = "year",
chains = 1, refresh = 2000, seed = 12345)
Fitting a univariate joint model.
Please note the warmup may be much slower than later iterations!
SAMPLING FOR MODEL 'jm' NOW (CHAIN 1).
Chain 1:
Chain 1: Gradient evaluation took 0.000242 seconds
Chain 1: 1000 transitions using 10 leapfrog steps per transition would take 2.42 seconds.
Chain 1: Adjust your expectations accordingly!
Chain 1:
Chain 1:
Chain 1: Iteration: 1 / 2000 [ 0%] (Warmup)
Chain 1: Iteration: 1001 / 2000 [ 50%] (Sampling)
Chain 1: Iteration: 2000 / 2000 [100%] (Sampling)
Chain 1:
Chain 1: Elapsed Time: 22.6376 seconds (Warmup)
Chain 1: 15.124 seconds (Sampling)
Chain 1: 37.7616 seconds (Total)
Chain 1:
The argument refresh = 2000
was specified so that Stan didn’t provide us with excessive progress updates whilst fitting the model. However, if you are fitting a model that will take several minutes or hours to fit, then you may wish to request progress updates quite regularly, for example setting refresh = 20
for every 20 iterations (by default the refresh argument is set to 1/10th of the total number of iterations).
The fitted model is returned as an object of the S3 class stanjm
. We have a variety of methods and postestimation functions available for this class, including: print
, summary
, plot
, fixef
, ranef
, coef
, VarCorr
, posterior_interval
, update
, and more. Here, we will examine the most basic output for the fitted joint model by typing print(mod1)
:
stan_jm
formula (Long1): logBili ~ sex + trt + year + (year  id)
family (Long1): gaussian [identity]
formula (Event): survival::Surv(futimeYears, death) ~ sex + trt
baseline hazard: bs
assoc: etavalue (Long1)

Longitudinal submodel: logBili
Median MAD_SD
(Intercept) 0.223 0.598
sexf 0.558 0.603
trt 0.166 0.407
year 0.210 0.038
sigma 0.353 0.017
Event submodel:
Median MAD_SD exp(Median)
(Intercept) 3.177 0.669 0.042
sexf 0.347 0.637 0.707
trt 0.708 0.448 0.493
Long1etavalue 1.343 0.245 3.830
bsplinescoef1 0.814 1.006 NA
bsplinescoef2 0.464 0.843 NA
bsplinescoef3 1.823 1.104 NA
bsplinescoef4 0.399 1.698 NA
bsplinescoef5 0.111 1.731 NA
bsplinescoef6 0.821 1.685 NA
Grouplevel error terms:
Groups Name Std.Dev. Corr
id Long1(Intercept) 1.2746
Long1year 0.1834 0.47
Num. levels: id 40
Sample avg. posterior predictive distribution
of longitudinal outcomes:
Median MAD_SD
Long1mean_PPD 0.585 0.029

For info on the priors used see help('prior_summary.stanreg').
The “Long1etavalue” row under “Event submodel” is our \(\alpha_{mq}\) parameter (\(m = 1\), \(q = 1\)). The estimated median of tells us that for each one unit increase in an individual’s underlying level of log serum bilirubin, their estimated log hazard of death increases by some amount. The mean absolute deviation (MAD) is provided as a more robust estimate of the standard deviation of the posterior distribution. In this case the MAD_SD for the association parameter indicates there is quite large uncertainty around the estimated association between log serum bilirubin and risk of death (recall this is a small dataset).
If we wanted some slightly more detailed output for each of the model parameters, as well as further details regarding the model estimation (for example computation time, number of longitudinal observations, number of individuals, type of baseline hazard, etc) we can instead use the summary
method:
Model Info:
function: stan_jm
formula (Long1): logBili ~ sex + trt + year + (year  id)
family (Long1): gaussian [identity]
formula (Event): survival::Surv(futimeYears, death) ~ sex + trt
baseline hazard: bs
assoc: etavalue (Long1)
algorithm: sampling
priors: see help('prior_summary')
sample: 1000 (posterior sample size)
num obs: 304 (Long1)
num subjects: 40
num events: 29 (72.5%)
groups: id (40)
runtime: 0.7 mins
Estimates:
mean sd 2.5%
Long1(Intercept) 0.225 0.602 0.970
Long1sexf 0.572 0.578 0.561
Long1trt 0.132 0.409 0.854
Long1year 0.213 0.039 0.145
Long1sigma 0.354 0.017 0.322
Long1mean_PPD 0.586 0.029 0.532
Event(Intercept) 3.183 0.682 4.583
Eventsexf 0.328 0.648 1.539
Eventtrt 0.740 0.459 1.661
Eventbsplinescoef1 0.925 1.071 3.410
Eventbsplinescoef2 0.452 0.843 1.283
Eventbsplinescoef3 1.840 1.196 4.357
Eventbsplinescoef4 0.374 1.653 2.982
Eventbsplinescoef5 0.126 1.714 3.490
Eventbsplinescoef6 1.095 1.847 5.386
AssocLong1etavalue 1.363 0.248 0.920
Sigma[id:Long1(Intercept),Long1(Intercept)] 1.625 0.390 1.031
Sigma[id:Long1year,Long1(Intercept)] 0.111 0.066 0.005
Sigma[id:Long1year,Long1year] 0.034 0.017 0.013
logposterior 329.723 10.559 351.610
97.5%
Long1(Intercept) 1.335
Long1sexf 1.644
Long1trt 0.719
Long1year 0.297
Long1sigma 0.389
Long1mean_PPD 0.644
Event(Intercept) 1.885
Eventsexf 1.003
Eventtrt 0.148
Eventbsplinescoef1 0.856
Eventbsplinescoef2 1.977
Eventbsplinescoef3 0.408
Eventbsplinescoef4 3.571
Eventbsplinescoef5 3.104
Eventbsplinescoef6 1.859
AssocLong1etavalue 1.898
Sigma[id:Long1(Intercept),Long1(Intercept)] 2.482
Sigma[id:Long1year,Long1(Intercept)] 0.258
Sigma[id:Long1year,Long1year] 0.076
logposterior 309.999
Diagnostics:
mcse Rhat n_eff
Long1(Intercept) 0.036 1.004 281
Long1sexf 0.032 1.004 324
Long1trt 0.035 1.000 134
Long1year 0.002 1.004 308
Long1sigma 0.001 1.000 764
Long1mean_PPD 0.001 1.002 944
Event(Intercept) 0.020 0.999 1216
Eventsexf 0.018 0.999 1336
Eventtrt 0.013 0.999 1179
Eventbsplinescoef1 0.044 1.006 595
Eventbsplinescoef2 0.028 0.999 883
Eventbsplinescoef3 0.051 1.001 558
Eventbsplinescoef4 0.077 1.001 460
Eventbsplinescoef5 0.076 1.003 514
Eventbsplinescoef6 0.064 1.000 827
AssocLong1etavalue 0.010 1.004 569
Sigma[id:Long1(Intercept),Long1(Intercept)] 0.031 1.018 158
Sigma[id:Long1year,Long1(Intercept)] 0.005 1.002 201
Sigma[id:Long1year,Long1year] 0.001 1.002 268
logposterior 0.858 1.004 151
For each parameter, mcse is Monte Carlo standard error, n_eff is a crude measure of effective sample size, and Rhat is the potential scale reduction factor on split chains (at convergence Rhat=1).
The easiest way to extract the correlation matrix for the random effects (aside from viewing the print
output) is to use the VarCorr
function (modelled on the VarCorr
function from the lme4 package). If you wish to extract the variances and covariances (instead of the standard deviations and correlations) then you can type the following to return a data frame with all of the relevant information:
grp var1 var2 vcov sdcor
1 id Long1(Intercept) <NA> 1.62465098 1.2746180
2 id Long1year <NA> 0.03362741 0.1833778
3 id Long1(Intercept) Long1year 0.11075635 0.4738511
In the previous example we were fitting a shared parameter joint model which assumed that the log hazard of the event (in this case the log hazard of death) at time t was linearly related to the subjectspecific expected value of the longitudinal marker (in this case the expected value of log serum bilirubin) also at time t. This is the default association structure, although it could be explicitly specified by setting the assoc = "etavalue"
argument.
However, let’s suppose we believe that the log hazard of death is actually related to both the current value of log serum bilirubin and the current rate of change in log serum bilirubin. To estimate this joint model we need to indicate that we want to also include the subjectspecific slope (at time t) from the longitudinal submodel as part of the association structure. We do this by setting the assoc
argument equal to a character vector c("etavalue", "etaslope")
which indicates our desired association structure:
mod2 < stan_jm(formulaLong = logBili ~ sex + trt + year + (year  id),
dataLong = pbcLong,
formulaEvent = survival::Surv(futimeYears, death) ~ sex + trt,
dataEvent = pbcSurv,
assoc = c("etavalue", "etaslope"),
time_var = "year",
chains = 1, refresh = 2000, seed = 12345)
In this example the subjectspecific slope is actually constant across time t since we have a linear trajectory. Note however that we could still use the "etaslope"
association structure even if we had a nonlinear subject specific trajectory (for example modelled using cubic splines or polynomials).
Suppose instead that we were interested in two repeatedly measured clinical biomarkers, log serum bilirubin and serum albumin, and their association with the risk of death. We may wish to model these two biomarkers, allowing for the correlation between them, and estimating their respective associations with the log hazard of death. We will fit a linear mixed effects submodel (identity link, normal distribution) for each biomarker with a patientspecific intercept and linear slope but no other covariates. In the event submodel we will include gender (sex
) and treatment (trt
) as baseline covariates. Each biomarker is assumed to be associated with the log hazard of death at time \(t\) via it’s expected value at time \(t\) (i.e. a current value association structure).
The model we are going to fit can therefore be specified as:
\[ y_{im}(t_{ijm}) \sim N(\mu_{im}(t_{ijm}), \sigma_m) \]
\[ \eta_{im}(t) = \mu_{im}(t) = \beta_{0m} + \beta_{1m} t + b_{0mi} + b_{1mi} t \]
\[ h_i(t) = h_0(t; \boldsymbol{\omega}) \exp(\gamma_1 w_{1i} + \gamma_2 w_{2i} + \alpha_{1i} \mu_{i1}(t) + \alpha_{2i} \mu_{i2}(t)) \]
where \(t\) is time in years, and \(w_{1i}\) and \(w_{2i}\) are, respectively, the gender and treatment indicators for individual \(i\).
(Note that due to the very small sample size, the clinical findings from this analysis should not to be overinterpreted!).
mod3 < stan_jm(
formulaLong = list(
logBili ~ sex + trt + year + (year  id),
albumin ~ sex + trt + year + (year  id)),
formulaEvent = survival::Surv(futimeYears, death) ~ sex + trt,
dataLong = pbcLong, dataEvent = pbcSurv,
time_var = "year",
chains = 1, refresh = 2000, seed = 12345)
Fitting a multivariate joint model.
Please note the warmup may be much slower than later iterations!
SAMPLING FOR MODEL 'jm' NOW (CHAIN 1).
Chain 1:
Chain 1: Gradient evaluation took 0.000404 seconds
Chain 1: 1000 transitions using 10 leapfrog steps per transition would take 4.04 seconds.
Chain 1: Adjust your expectations accordingly!
Chain 1:
Chain 1:
Chain 1: Iteration: 1 / 2000 [ 0%] (Warmup)
Chain 1: Iteration: 1001 / 2000 [ 50%] (Sampling)
Chain 1: Iteration: 2000 / 2000 [100%] (Sampling)
Chain 1:
Chain 1: Elapsed Time: 38.4462 seconds (Warmup)
Chain 1: 35.1984 seconds (Sampling)
Chain 1: 73.6447 seconds (Total)
Chain 1:
We can now examine the output from the fitted model, for example
stan_jm
formula (Long1): logBili ~ sex + trt + year + (year  id)
family (Long1): gaussian [identity]
formula (Long2): albumin ~ sex + trt + year + (year  id)
family (Long2): gaussian [identity]
formula (Event): survival::Surv(futimeYears, death) ~ sex + trt
baseline hazard: bs
assoc: etavalue (Long1), etavalue (Long2)

Longitudinal submodel 1: logBili
Median MAD_SD
(Intercept) 0.151 0.634
sexf 0.577 0.584
trt 0.032 0.391
year 0.224 0.041
sigma 0.355 0.016
Longitudinal submodel 2: albumin
Median MAD_SD
(Intercept) 3.478 0.255
sexf 0.027 0.234
trt 0.005 0.167
year 0.156 0.024
sigma 0.291 0.013
Event submodel:
Median MAD_SD exp(Median)
(Intercept) 6.770 3.065 871.592
sexf 0.118 0.703 0.888
trt 0.488 0.523 0.614
Long1etavalue 0.811 0.274 2.250
Long2etavalue 3.090 0.915 0.045
bsplinescoef1 1.026 1.144 NA
bsplinescoef2 0.546 0.946 NA
bsplinescoef3 2.713 1.337 NA
bsplinescoef4 0.549 1.762 NA
bsplinescoef5 1.297 1.717 NA
bsplinescoef6 2.614 1.856 NA
Grouplevel error terms:
Groups Name Std.Dev. Corr
id Long1(Intercept) 1.25487
Long1year 0.18745 0.47
Long2(Intercept) 0.52113 0.65 0.50
Long2year 0.09534 0.55 0.80 0.44
Num. levels: id 40
Sample avg. posterior predictive distribution
of longitudinal outcomes:
Median MAD_SD
Long1mean_PPD 0.586 0.028
Long2mean_PPD 3.345 0.023

For info on the priors used see help('prior_summary.stanreg').
or we can examine the summary output for the association parameters alone:
Model Info:
function: stan_jm
formula (Long1): logBili ~ sex + trt + year + (year  id)
family (Long1): gaussian [identity]
formula (Long2): albumin ~ sex + trt + year + (year  id)
family (Long2): gaussian [identity]
formula (Event): survival::Surv(futimeYears, death) ~ sex + trt
baseline hazard: bs
assoc: etavalue (Long1), etavalue (Long2)
algorithm: sampling
priors: see help('prior_summary')
sample: 1000 (posterior sample size)
num obs: 304 (Long1), 304 (Long2)
num subjects: 40
num events: 29 (72.5%)
groups: id (40)
runtime: 1.2 mins
Estimates:
mean sd 2.5% 25% 50% 75% 97.5%
AssocLong1etavalue 0.826 0.285 0.278 0.641 0.811 1.009 1.409
AssocLong2etavalue 3.149 0.929 5.175 3.729 3.090 2.499 1.556
Diagnostics:
mcse Rhat n_eff
AssocLong1etavalue 0.011 1.000 723
AssocLong2etavalue 0.039 0.999 573
For each parameter, mcse is Monte Carlo standard error, n_eff is a crude measure of effective sample size, and Rhat is the potential scale reduction factor on split chains (at convergence Rhat=1).
We can also access the range of postestimation functions (described in the stan_jm
and related help documentation; see for example help(posterior_traj)
or help(posterior_survfit)
).
Predicted individualspecific biomarker values can be obtained using either the posterior_traj
or posterior_predict
function. The posterior_traj
is preferable, because it can be used to obtain the biomarker values at a series of evenly spaced time points between baseline and the individual’s event or censoring time by using the default interpolate = TRUE
option. Whereas, the posterior_predict
function only provides the predicted biomarker values at the observed time points, or the time points in the new data. Predicting the biomarker values at a series of evenly spaced time points can be convenient because they can be easily used for plotting the longitudinal trajectory. Moreover, by default the posterior_traj
returns a data frame with variables corresponding to the individual ID, the time, the predicted mean biomarker value, the limits for the 95% credible interval (i.e. uncertainty interval for the predicted mean biomarker value), and limits for the 95% prediction interval (i.e. uncertainty interval for a predicted biomarker data point), where the level for the uncertainty intervals can be changed via the prob
argument. Conversely, the posterior_predict
function returns an \(S\) by \(N\) matrix of predictions where \(S\) is the number of posterior draws and \(N\) is the number of prediction time points (note that this return type can also be obtained for posterior_traj
by specifying the argument return_matrix = TRUE
).
As an example, let’s plot the predicted individualspecific longitudinal trajectories for each of the two biomarkers (log serum bilirubin and serum albumin) in the multivariate joint model estimated above. We will do this for three individuals (IDs 6, 7 and 8) who were included in the model estimation.
Here are the plots for log serum bilirubin:
and here are the plots for serum albumin:
The m
argument specifies which biomarker we want to predict for (only relevant for a multivariate joint model). The ids
argument is optional, and specifies a subset of individuals for whom we want to predict. In the plotting method, the plot_observed = TRUE
specifies that we want to include the observed biomarker values in the plot of the longitudinal trajectory.
If we wanted to extrapolate the trajectory forward from the event or censoring time for each individual, then this can be easily achieved by specifying extrapolate = TRUE
in the posterior_traj
call. For example, here is the plot for log serum bilirubin with extrapolation:
p3 < posterior_traj(mod3, m = 1, ids = 6:8, extrapolate = TRUE)
pp3 < plot(p3, plot_observed = TRUE, vline = TRUE)
pp3
and for serum albumin with extrapolation:
p4 < posterior_traj(mod3, m = 2, ids = 6:8, extrapolate = TRUE)
pp4 < plot(p4, plot_observed = TRUE, vline = TRUE)
pp4
Here, we included the vline = TRUE
which adds a vertical dashed line at the timing of the individual’s event or censoring time. The interpolation and extrapolation of the biomarker trajectory can be further controlled through the control
argument to the posterior_traj
function; for example, we could specify the number of time points at which to predict, the distance by which to extrapolate, and so on.
We could customize these plots further, for example, by using any of the ggplot2 functionality or using the additional arguments described in help(plot.predict.stanjm)
.
Predicted individualspecific survival probabilities and/or survival curves can be obtained using the posterior_survfit
function. The function by default returns a data frame with the individual ID, the time, and the predicted survival probability (posterior mean and limits for the 95% credible interval). The uncertainty level for the credible interval can be changed via the prob
argument. By default, individualspecific survival probabilities are calculated conditional on the individual’s last known survival time. When we are predicting survival probabilities for individuals that were used in the estimation of the model (i.e. insample individuals, where no new covariate data is provided), then the individual’s “last known survival time” will be their event or censoring time. (Note that if we wanted didn’t want to condition on the individual’s last known survival time, then we could specify condition = FALSE
, but we probably wouldn’t want to do this unless we were calculating marginal or standardised survival probabilities, which are discussed later).
The default argument extrapolate = TRUE
specifies that the individualspecific conditional survival probabilities will be calculated at evenly spaced time points between the individual’s last known survival time and the maximum follow up time that was observed in the estimation sample. The behaviour of the extrapolation can be further controlled via the control
argument. If we were to specify extrapolate = FALSE
then the survival probabilities would only be calculated at one time point, which could be specified in the times
argument (or otherwise would default to the individual’s last known survival time).
As an example, let plot the predicted individualspecific conditional survival curve for the same three individual’s that were used in the previous example. The predicted survival curve will be obtained under the multivariate joint model estimated above.
We could customize the plot further, for example, by using any of the ggplot2 functionality or using the additional arguments described in help(plot.survfit.stanjm)
.
The package also provides a convenience plotting function, which combines plots of the individualspecific longitudinal trajectories, and the individualspecific survival function. We can demonstrate this by replotting the predictions for the three individuals in the previous example:
Here we can see the strong relationship between the underlying values of the biomarkers and mortality. Patient 8
who, relative to patients 6
and 7
, has a higher underlying value for log serum bilirubin and a lower underlying value for serum albumin at the end of their follow up has a far worse predicted probability of survival.
Let us take an individual from our training data, in this case the individual with subject ID value 8
. However, we will pretend this individual was not a member of our training data and rather that they are a new individual for whom we have obtained new biomarker measurements. Our goal is to obtain predictions for the longitudinal trajectory for this individual, and their conditional survival curve, given that we know they are conditional on their biomarker measurements we currently have available.
First, let’s extract the data for subject 8
and then rename their subject ID value so that they appear to be an individual who was not included in our training dataset:
ndL < pbcLong[pbcLong$id == 8, , drop = FALSE]
ndE < pbcSurv[pbcSurv$id == 8, , drop = FALSE]
ndL$id < paste0("new_patient")
ndE$id < paste0("new_patient")
Note that we have both the longitudinal data and event data for this new individual. We require data for both submodels because we are going to generate dynamic predictions that require drawing new individualspecific parameters (i.e. random effects) for this individual conditional on their observed data. That means we need to evaluate the likelihood for the full joint model and that requires both the longitudinal and event data (note however that the status indicator death
will be ignored, since it is assumed that the individual we are predicting for is still alive at the time we wish to generate the predictions).
Now we can pass this data to the posterior_traj
function in the same way as for the insample individuals, except we will now specify the newdataLong
and newdataEvent
arguments. We will also specify the last_time
argument so that the function knows which variable in the event data specifies the individual’s last known survival time (the default behaviour is to use the time of the last biomarker measurement).
Our predictions for this new individual for the log serum bilirubin trajectory can be obtained using:
Drawing new random effects for 1 individuals. Monitoring progress:

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and for the serum albumin trajectory:
Drawing new random effects for 1 individuals. Monitoring progress:

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For the conditional survival probabilities we use similar information, provided to the posterior_survfit
function:
Drawing new random effects for 1 individuals. Monitoring progress:

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We can then use the plot_stack_jm
function, as we saw in a previous example, to stack the plots of the longitudinal trajectory and the conditional survival curve:
Here we see that the predicted longitudinal trajectories and conditional survival curve for this individual, obtained using the dynamic predictions approach, are similar to the predictions we obtained when we used their individualspecific parameters from the original model estimation. This is because in both situations we are conditioning on the same outcome data.
Side note: We can even compare the estimated individual specific parameters obtained under the two approaches. For example, here is the posterior mean for the estimated individualspecific parameters for individual 8
from the fitted model:
$`(Intercept)`
[1] 1.70357
$year
[1] 0.1101605
$`(Intercept)`
[1] 0.4906283
$year
[1] 0.04774153
and here is the mean of the draws for the individualspecific parameters for individual 8
under the dynamic predictions approach:
b[Long1(Intercept) id:new_patient] b[Long1year id:new_patient]
1.74039437 0.11208411
b[Long2(Intercept) id:new_patient] b[Long2year id:new_patient]
0.47535238 0.04009939
Suppose we wanted to predict the longitudinal trajectory for each of the biomarkers, marginalising over the distribution of the individualspecific parameters. To do this, we can pass a new data frame with the covariate values we want to use in the predictions. Here, we will demonstrate this by obtaining the predicted trajectory for log serum bilirubin, under the multivariate joint model that was estimated previously. Our prediction data will require the variables year
, sex
and trt
, since these were the covariates used in the longitudinal submodel.
We will predict the value of log serum bilirubin at years 0 through 10, for each combination of sex
and trt
. We also need to include the id
variable in our prediction data because this is relevant to the longitudinal submodel. Since we want to marginalise over the individualspecific parameters (i.e. individuallevel random effects) we need to note two things:
First, the values for the id
variable must not match any individual used in the model estimation. Here, we use the following id
values: "male_notrt"
, "female_notrt"
, "male_trt"
, and "female_trt"
, since each individual in our prediction data represents a different combination of sex
and trt
. However, we could have given the individuals any id
value just as long as is didn’t match an individual who was used in the model estimation
Second, we need to specify the argument dynamic = FALSE
when calling posterior_traj
. This specifies that we do not want to draw new individualspecific parameters conditional on outcome data observed up to some time \(t\). Instead, we want predictions that marginalise over the distribution of individualspecific parameters and are therefore conditional only on covariates and not conditional on outcome data for the new individuals.
Here is our prediction data:
ndL < expand.grid(year = seq(0, 10, 1),
sex = c("m", "f"),
trt = 0:1)
ndL$id < rep(c("male_notrt", "female_notrt",
"male_trt", "female_trt"), each = 11)
ndL < ndL[, c(4,1,2,3)]
str(ndL)
'data.frame': 44 obs. of 4 variables:
$ id : chr "male_notrt" "male_notrt" "male_notrt" "male_notrt" ...
$ year: num 0 1 2 3 4 5 6 7 8 9 ...
$ sex : Factor w/ 2 levels "m","f": 1 1 1 1 1 1 1 1 1 1 ...
$ trt : int 0 0 0 0 0 0 0 0 0 0 ...
And to predict the marginal longitudinal trajectory for log serum bilirubin under each covariate profile and plot it we can type:
p1 < posterior_traj(mod3, m = 1, newdataLong = ndL, dynamic = FALSE)
plot(p1) + ggplot2::coord_cartesian(ylim = c(10,15))
Because we are marginalising over the distribution of the individualspecific parameters, we are incorporating all the variation related to betweenindividual differences, and therefore the prediction interval is wide (shown by the shaded area around the marginal longitudinal trajectory). The magnitude of the effects of both sex
and trt
are relatively small compared to the populationlevel effect of year
and the betweenindividual variation in the intercept and slope. For example, here are the point estimates for the populationlevel effects of sex
, trt
, and year
:
(Intercept) sexf trt year
0.15106619 0.57729068 0.03177512 0.22438924
and here are the standard deviations for the individuallevel random effects:
Groups Name Std.Dev. Corr
id Long1(Intercept) 1.25487
Long1year 0.18745 0.474
Long2(Intercept) 0.52113 0.646 0.499
Long2year 0.09534 0.553 0.801 0.442
This shows us that the point estimates for the populationlevel effects of sex
and trt
are 0.57 and 0.10, respectively, whereas the standard deviation for the individualspecific intercept and slope parameters are 1.24 and 0.19; hence, any differences due to the populationlevel effects of gender and treatment (i.e. differences in the black line across the four panels of the plot) are swamped by the width of the uncertainty intervals (i.e. the grey shaded areas).
In this example we show how a standardised survival curve can be obtained, where the \(i = 1,...,N^{pred}\) individuals used in generating the standardised survival curve are the same individuals that were used in estimating the model. We will obtain the survival curve for the multivariate joint model estimated in an earlier example (mod3
). The standardise = TRUE
argument to posterior_survfit
specifies that we want to obtain individualspecific predictions of the survival curve and then average these. Because, in practical terms, we need to obtain survival probabilities at time \(t\) for each individual and then average them we want to explicitly specify the values of \(t\) we want to use (and the same values of \(t\) will be used for individuals). We specify the values of \(t\) to use via the times
argument; here we will predict the standardised survival curve at time 0 and then for convenience we can just specify extrapolate = TRUE
(which is the default anyway) which will mean we automatically predict at 10 evenly spaced time points between 0 and the maximum event or censoring time.
p1 < posterior_survfit(mod3, standardise = TRUE, times = 0)
head(p1) # data frame with standardised survival probabilities
year survpred ci_lb ci_ub
1 0.0000 1.0000 1.0000 1.0000
2 1.0154 0.8301 0.7656 0.9044
3 2.0307 0.7385 0.6684 0.7880
4 3.0461 0.6863 0.6223 0.7383
5 4.0614 0.6411 0.5848 0.6946
6 5.0768 0.5984 0.5403 0.6477
By “true” value of the biomarker, we mean the value of the biomarker which is not subject to measurement error or discrete time observation. Of course, for the expected value from the longitudinal submodel to be considered the socalled “true” underlying biomarker value, we would need to have specified the longitudinal submodel appropriately!↩
We refer the reader to the priors vignette for a discussion of the possible prior distributions.↩
These random draws from the posterior distribution of the groupspecific parameters are stored each time a joint model is estimated using stan_glmer
, stan_mvmer
, or stan_jm
; they are saved under an ID value called "_NEW_"
↩