# Joint Modelling of Functional Regression and FPCA

#### 2026-04-28

# Introduction

This vignette describes the joint FPCA option that is available across the `refundBayes` regression functions:

- `sofr_bayes()` – Bayesian Scalar-on-Function Regression,
- `fcox_bayes()` – Bayesian Functional Cox Regression,
- `fofr_bayes()` – Bayesian Function-on-Function Regression.

In every case, the option is exposed via the same `joint_FPCA` argument: a logical (`TRUE` / `FALSE`) vector of length equal to the number of functional predictors. When the \(i\)-th entry is `TRUE`, the observed functional predictor \(W_i(s)\) is not used directly as a covariate; instead, the regression model is built on top of a functional principal component analysis (FPCA) sub-model for \(W_i(s)\), and the FPC scores are sampled jointly with the regression coefficients.

The joint-FPCA option replaces each functional predictor \(W_i(s)\) with a low-rank FPCA representation \(\sum_{j=1}^{J}\xi_{ij}\,\phi_j(s)\) — fixed eigenfunctions \(\phi_j\) from `refund::fpca.sc()`, subject-specific scores \(\xi_{ij}\) centered on their frequentist FPCA estimates — and samples those scores jointly with the regression coefficient \(\beta(\cdot)\), propagating predictor measurement-error uncertainty into the posterior of \(\beta\) and correcting the errors-in-variables attenuation that arises when noisy \(W_i\) is plugged into the regression as if it were observed exactly.

The methodology mirrors Section 4 of Jiang, Crainiceanu, and Cui (2025), Tutorial on Bayesian Functional Regression Using Stan, Statistics in Medicine 44(20–22), e70265. The default option is (`joint_FPCA = NULL`) which do not adopt joint modelling.

# Why Joint FPCA?

In standard scalar-on-function or functional Cox regression, the integral \(\int W_i(s)\beta(s)\,ds\) is plugged in as if \(W_i(s)\) were observed without error. In practice, the curves are sampled at finitely many points and are typically corrupted by measurement noise. Treating \(W_i(s)\) as exact ignores this uncertainty and biases the regression coefficient \(\beta(\cdot)\) toward zero (the classical errors-in-variables attenuation), and underestimates the posterior credible-interval width.

The joint FPCA approach replaces \(W_i(s)\) by a low-rank FPCA representation

\[W_i(s) \;=\; \mu(s) \;+\; \sum_{j=1}^{J} \xi_{ij}\,\phi_j(s) \;+\; \epsilon_i(s),\]

and treats the subject-specific scores \(\xi_{ij}\) as parameters that are sampled together with the regression coefficients. The FPC eigenfunctions \(\phi_j(s)\) are estimated once with `refund::fpca.sc()` and held fixed, and the initial frequentist FPCA scores \(\hat\xi_{ij}\) are used as the prior mean for \(\xi_{ij}\). This propagates the FPCA uncertainty into the posterior of \(\beta(\cdot)\), so the regression credible bands are correctly inflated.

# The Joint Bayesian Model We Adopted

The model has two layers that share the FPC scores \(\xi_{ij}\):

1. an FPCA likelihood on the observed functional predictor; and
2. the outcome regression (SoFR / FCox / FoFR), in which the scores \(\xi_{ij}\) enter the linear predictor.

Throughout this section we assume one functional predictor for clarity; the multi-predictor case is identical with one set of FPCA quantities per functional predictor.

## FPCA likelihood

For each subject \(i = 1, \ldots, n\) and each grid point \(s_m\), \(m = 1, \ldots, M\),

\[W_i(s_m) \;=\; \sum_{j=1}^{J} \xi_{ij}\,\phi_j(s_m) \;+\; \epsilon_i(s_m), \qquad \epsilon_i(s_m) \stackrel{\text{iid}}{\sim} N\!\left(0, \sigma_e^2\right),\]

so that

\[\log p(\mathbf{W} \mid \boldsymbol\xi, \boldsymbol\Phi, \sigma_e) \;=\; -\, n\,M\,\log\sigma_e \;-\; \frac{1}{2\sigma_e^2}\sum_{i=1}^{n}\sum_{m=1}^{M}\!\left\{W_i(s_m) - \sum_{j=1}^{J}\xi_{ij}\,\phi_j(s_m)\right\}^2.\]

In Stan code, this corresponds to

    target += - N_num * M_num_i * log(sigma_e_i)
              - sum((xi_i * Phi_mat_i - M_mat_i)^2) / (2 * sigma_e_i^2);

Here `M_mat_i` is the observed \(n \times M\) matrix of \(W_i(s_m)\) values, `Phi_mat_i` is the \(J \times M\) matrix of fixed eigenfunctions, and `xi_i` is the \(n \times J\) matrix of FPC score parameters.

## FPC score prior

Each FPC score is given an independent Gaussian prior centered on the initial frequentist FPCA score with eigenvalue-determined scale,

\[\xi_{ij} \;\sim\; N\!\left(\hat\xi_{ij},\,\lambda_j^{2}\right), \qquad i = 1,\ldots,n,\;\; j = 1,\ldots,J.\]

In Stan code, the kernel is implemented as

    for (nj in 1:J_num_i) {
       target += - N_num * log(lambda_i[nj])
                 - sum((xi_i[, nj] - xi_hat_i[, nj])^2) / (2 * lambda_i[nj]^2);
       target += inv_gamma_lpdf(lambda_i[nj]^2 | 0.001, 0.001);
    }
    target += inv_gamma_lpdf(sigma_e_i^2 | 0.001, 0.001);

The eigenvalue scales \(\lambda_j^2\) and the residual scale \(\sigma_e^2\) both receive weakly informative inverse-Gamma priors. The prior mean \(\hat\xi_{ij}\) is computed once via `refund::fpca.sc()` on the observed functional data.

## Plugging the FPCA into the linear predictor

Because \(W_i(s) = \sum_j \xi_{ij}\,\phi_j(s)\) in the FPCA representation, the functional integral that drives every regression model in `refundBayes` becomes

\[\int_{\mathcal{S}} W_i(s)\,\beta(s)\,ds \;=\; \sum_{j=1}^{J}\xi_{ij}\,\underbrace{\int_{\mathcal{S}} \phi_j(s)\,\beta(s)\,ds}_{\,\equiv\,\alpha_j}.\]

Expanding \(\beta(s)\) in the spline basis \(\beta(s) = \sum_{k=1}^{K} b_k\,\psi_k(s)\), we have

\[\alpha_j \;=\; \sum_{k=1}^{K} b_k \int_{\mathcal{S}} \phi_j(s)\,\psi_k(s)\,ds \;\equiv\; \sum_{k=1}^{K} \mathbf{X}_{jk}\,b_k,\]

where the \(J \times K\) FPCA-spline cross-product matrix \(\mathbf{X}\) is approximated by the Riemann sum

\[\mathbf{X}_{jk} \;\approx\; \sum_{m=1}^{M} L_m\,\phi_j(s_m)\,\psi_k(s_m)\]

using the same integration weights \(L_m\) as in the standard regression formulation. After the spectral reparameterization of \(\mathbf{X}\) via \(\mathbf{S} = \int \boldsymbol\psi''(s)\boldsymbol\psi''(s)^t\,ds\) (the penalty matrix), the design matrix \(\mathbf{X}\) is split into a penalized random-effect block \(\tilde{\mathbf{X}}_r\) (size \(K_r \times J\)) and an unpenalized fixed-effect block \(\tilde{\mathbf{X}}_f\) (size \(K_f \times J\)). The linear predictor for SoFR / FCox is then

\[\eta_i \;=\; \eta_0 \;+\; \mathbf{Z}_i^{t}\boldsymbol\gamma \;+\; \boldsymbol\xi_i^{t}\!\left(\tilde{\mathbf{X}}_r^{t}\,\tilde{\mathbf{b}}_r \;+\; \tilde{\mathbf{X}}_f^{t}\,\tilde{\mathbf{b}}_f\right),\]

with \(\tilde{\mathbf{b}}_r \sim N(\mathbf{0},\sigma_b^2\mathbf{I})\) and \(\tilde{\mathbf{b}}_f\) assigned non-informative priors — exactly the same prior structure used for the regression coefficients in the no-joint-FPCA case. This is the chunk in the Stan code:

    mu += xi_i * (X_mat_r_i' * br_i + X_mat_f_i' * bf_i);

For the function-on-function regression model the integration over \(s\) is combined with a response-domain expansion in the basis \(\boldsymbol\psi(t)\) of size \(K_{\text{resp}}\), so that the bivariate coefficient \(\beta(s,t) = \sum_{k=1}^{K_{\text{resp}}}\beta_{k}(s)\,\psi_k(t)\) leads to

\[\int W_i(s)\,\beta(s,t)\,ds \;=\; \boldsymbol\xi_i^{t}\!\left(\tilde{\mathbf{X}}_r^{t}\,\mathbf{B}_r \;+\; \tilde{\mathbf{X}}_f^{t}\,\mathbf{B}_f\right)\boldsymbol\psi(t),\]

where now \(\mathbf{B}_r\) is a \(K_r \times K_{\text{resp}}\) matrix and \(\mathbf{B}_f\) is a \(K_f \times K_{\text{resp}}\) matrix of bivariate coefficients. The resulting Stan model contribution is

    mu += (xi_i * (X_mat_r_i' * br_i + X_mat_f_i' * bf_i)) * Psi_mat;

This is the only structural difference relative to the joint-FPCA SoFR / FCox model: the spline coefficients become matrices indexed by both the predictor-domain and the response-domain bases, and the same row-wise response-domain penalty as in the standard FoFR is applied.

## Full Bayesian model

The complete joint model (one functional predictor) reads

\[\begin{cases} \textbf{Outcome regression}\colon\quad \eta_i = \eta_0 + \mathbf{Z}_i^{t}\boldsymbol\gamma + \boldsymbol\xi_i^{t}\!\left(\tilde{\mathbf{X}}_r^{t}\,\tilde{\mathbf{b}}_r + \tilde{\mathbf{X}}_f^{t}\,\tilde{\mathbf{b}}_f\right) \\[6pt] \textbf{FPCA likelihood}\colon\quad W_i(s_m) \mid \boldsymbol\xi_i, \boldsymbol\Phi, \sigma_e \sim N\!\left(\sum_j \xi_{ij}\phi_j(s_m),\,\sigma_e^2\right) \\[4pt] \textbf{FPC score prior}\colon\quad \xi_{ij} \sim N(\hat\xi_{ij},\,\lambda_j^{2}) \\[4pt] \textbf{Hyperpriors}\colon\quad \tilde{\mathbf{b}}_r \sim N(\mathbf{0},\sigma_b^2\mathbf{I}),\;\; \sigma_b^2,\,\sigma_e^2,\,\lambda_j^{2} \sim \text{Inv-Gamma}(0.001, 0.001) \\[4pt] \textbf{Outcome distribution}\colon\quad Y_i \sim p(Y_i \mid \eta_i)\quad\text{(Gaussian / binomial / Cox / functional)} \end{cases}\]

The four building blocks (outcome regression, FPCA likelihood, score prior, hyperpriors) are independent of the family; only the outcome distribution in the last line changes. This is precisely why the joint FPCA option is available across SoFR, FCox, and FoFR with the same `joint_FPCA` argument.

## Note on the prior mean of \(\xi_{ij}\)

The prior is centered on the initial frequentist FPCA score \(\hat\xi_{ij}\) returned by `refund::fpca.sc()`, not on \(0\). This corresponds to using the standard FPCA fit as a working informative prior. As \(\lambda_j^{2}\) becomes large the prior reverts to a diffuse Gaussian and the joint model reduces (in spirit) to plugging in the sample FPC scores; for moderate \(\lambda_j^{2}\) the posterior trades off the FPCA fit and the regression likelihood. The observed-data matrix \(W_i(s_m)\) is passed to Stan in its original (uncentered) form, exactly as in Section 4 of the Tutorial supplement.

## Identifying the functional data and the integration weights

The joint FPCA design matrix \(\mathbf{X}_{jk}\) is built with the same Riemann-sum integration weights as the no-joint-FPCA design matrix. Specifically, the convention is:

- in `s(tindex, by = lmat * wmat, ...)`, the rightmost variable in the product (here `wmat`) is treated as the observed functional data \(W_i(\cdot)\);
- the product of the remaining variables (here `lmat`) is treated as the Riemann-sum integration weight matrix.

This matches the Tutorial supplementary code and means that you do not need to change your formulas when turning joint FPCA on or off; only the `joint_FPCA` argument changes.

# How to enable joint FPCA

Every regression function in `refundBayes` accepts the same argument:

    joint_FPCA = c(TRUE, FALSE, ...)   # one entry per s(...) functional term

The default is `joint_FPCA = NULL`, which is treated as `rep(FALSE, n_func)` and gives exactly the no-joint-FPCA behavior of the respective regression vignettes. Below we walk through one example for each of `sofr_bayes()`, `fcox_bayes()`, and `fofr_bayes()`.

# Example 1: Joint FPCA in `sofr_bayes()`

We reuse the simulation setup from the SoFR vignette and switch on joint FPCA for the (single) functional predictor.

## Simulate Data

    set.seed(123)
    n <- 100
    M <- 50
    tgrid <- seq(0, 1, length.out = M)
    dt    <- tgrid[2] - tgrid[1]
    tmat  <- matrix(rep(tgrid, each = n), nrow = n)
    lmat  <- matrix(dt, nrow = n, ncol = M)

    # Smooth latent trajectory + measurement noise on the functional predictor
    D_true <- t(apply(matrix(rnorm(n * M), n, M), 1, cumsum)) / sqrt(M)
    wmat   <- D_true + matrix(rnorm(n * M, sd = 0.3), nrow = n) ## noisy

    beta_true <- sin(2 * pi * tgrid)
    X1        <- rnorm(n)
    eta       <- 0.5 * X1 + D_true %*% (beta_true * dt)   ## regression on D, not W
    prob      <- plogis(eta)
    y         <- rbinom(n, 1, prob)

    data.SoFR <- data.frame(y = y, X1 = X1)
    data.SoFR$tmat <- tmat
    data.SoFR$lmat <- lmat
    data.SoFR$wmat <- wmat

Notice that the outcome is generated from the latent trajectory \(D_i(s)\) but only the noisy \(W_i(s) = D_i(s) + \epsilon_i(s)\) is observed. This is exactly the setting that motivates joint FPCA.

## Fit the joint FPCA SoFR model

    library(refundBayes)

    fit_sofr_joint <- sofr_bayes(
      formula    = y ~ X1 + s(tmat, by = lmat * wmat, bs = "cc", k = 10),
      data       = data.SoFR,
      family     = binomial(),
      joint_FPCA = c(TRUE),       ## << turn joint FPCA on for the (only) functional term
      niter      = 1500,
      nwarmup    = 500,
      nchain     = 3,
      ncores     = 3
    )

For comparison, the no-joint-FPCA fit uses the same call without the `joint_FPCA` argument:

    fit_sofr_plain <- sofr_bayes(
      formula = y ~ X1 + s(tmat, by = lmat * wmat, bs = "cc", k = 10),
      data    = data.SoFR,
      family  = binomial(),
      niter   = 1500, nwarmup = 500, nchain = 3, ncores = 3
    )

The two posterior estimates of \(\beta(s)\) can be plotted together:

    plot(fit_sofr_joint)
    plot(fit_sofr_plain)

Joint-FPCA credible bands are typically wider in the regions where the measurement noise on the functional predictor is informative.

## Joint-FPCA quantities exposed in the Stan fit

When `joint_FPCA = TRUE` for the \(i\)-th functional predictor, the Stan fit additionally exposes:

Stan parameter | Meaning
--- | ---
`xi_i` | \(n \times J\) matrix of joint FPC scores
`lambda_i` | \(J\)-vector of FPC eigenvalue SDs
`sigma_e_i` | scalar, FPC residual SD

These can be inspected with `rstan::extract()` for further analysis (for example, plotting the posterior of the FPC scores or the eigenvalue SDs).

# Example 2: Joint FPCA in `fcox_bayes()`

The functional Cox setup is identical to the SoFR setup, with one extra piece (the censoring vector). The joint FPCA model is wired in by setting `joint_FPCA = c(TRUE)` on a model that has one functional predictor:

    library(refundBayes)

    ## Use the example dataset shipped with the FCox vignette
    Func_Cox_Data <- readRDS("data/example_data_Cox.rds")
    Func_Cox_Data$wmat <- Func_Cox_Data$MIMS
    Func_Cox_Data$cens <- 1 - Func_Cox_Data$event

    fit_cox_joint <- fcox_bayes(
      formula    = survtime ~ X1 + s(tmat, by = lmat * wmat, bs = "cc", k = 10),
      data       = Func_Cox_Data,
      cens       = Func_Cox_Data$cens,
      joint_FPCA = c(TRUE),
      niter      = 5000,
      nwarmup    = 2000,
      nchain     = 1,
      ncores     = 1
    )

The Stan code generated under the hood matches Section 4 of the Tutorial supplement: the linear predictor

\[\eta_i \;=\; \mathbf{Z}_i^{t}\boldsymbol\gamma + \boldsymbol\xi_i^{t}\!\left(\tilde{\mathbf{X}}_r^{t}\tilde{\mathbf{b}}_r + \tilde{\mathbf{X}}_f^{t}\tilde{\mathbf{b}}_f\right)\]

is plugged into the Cox log-likelihood

\[\ell(\mathbf{Y},\boldsymbol\delta) \;=\; \sum_{i=1}^{n}\left[(1-\delta_i)\!\left\{\log h_0(Y_i) + \eta_i - H_0(Y_i)e^{\eta_i}\right\} + \delta_i\!\left\{-H_0(Y_i)e^{\eta_i}\right\}\right],\]

with the same M-spline / I-spline baseline-hazard construction as in the no-joint case. The FPCA likelihood and the FPC-score prior are appended to the joint Stan target function exactly as described in The Joint Bayesian Model We Adopted above.

## Inspecting the generated Stan code

    fcox_code <- fcox_bayes(
      formula    = survtime ~ X1 + s(tmat, by = lmat * wmat, bs = "cc", k = 10),
      data       = Func_Cox_Data,
      cens       = Func_Cox_Data$cens,
      joint_FPCA = c(TRUE),
      runStan    = FALSE
    )
    cat(fcox_code$stancode)

You will see the FPCA-related declarations in the `data` block (`Phi_mat_1`, `xi_hat_1`, `M_mat_1`, `J_num_1`, `M_num_1`, `X_mat_r_1`, `X_mat_f_1`), the FPCA-related parameters in the `parameters` block (`xi_1`, `lambda_1`, `sigma_e_1`), and the FPCA log-likelihood + score prior contributions in the `model` block.

# Example 3: Joint FPCA in `fofr_bayes()`

For function-on-function regression, joint FPCA on the functional predictor is constructed in exactly the same way — only the contribution to \(\boldsymbol\mu_i(t)\) now carries the response-domain basis \(\boldsymbol\Psi\):

\[\int W_i(s)\,\beta(s,t)\,ds \;=\; \boldsymbol\xi_i^{t}\!\left(\tilde{\mathbf{X}}_r^{t}\,\mathbf{B}_r + \tilde{\mathbf{X}}_f^{t}\,\mathbf{B}_f\right)\boldsymbol\psi(t),\]

with \(\mathbf{B}_r,\mathbf{B}_f\) as matrices of size \(K_r \times K_{\text{resp}}\) and \(K_f \times K_{\text{resp}}\) respectively. Both directions of smoothness (\(s\) via the random-effect reparameterization and \(t\) via the response-domain penalty) are imposed exactly as in the no-joint FoFR model.

## Simulated FoFR with measurement error on the predictor

    library(refundBayes)
    set.seed(42)

    n <- 200
    L <- 30
    M <- 30
    sindex <- seq(0, 1, length.out = L)
    tindex <- seq(0, 1, length.out = M)

    # Smooth latent functional predictor + measurement noise
    D_true <- matrix(0, nrow = n, ncol = L)
    for (i in 1:n) {
      D_true[i, ] <- rnorm(1) * sin(2 * pi * sindex) +
                     rnorm(1) * cos(2 * pi * sindex) +
                     rnorm(1) * sin(4 * pi * sindex)
    }
    X_func <- D_true + matrix(rnorm(n * L, sd = 0.3), nrow = n)   ## noisy

    age <- rnorm(n)

    # True coefficient functions
    beta_true  <- outer(sin(2 * pi * sindex), cos(2 * pi * tindex))
    alpha_true <- 0.5 * sin(pi * tindex)

    # Generate response from the latent D_true (not from X_func!)
    signal_scalar <- outer(age, alpha_true)
    signal_func   <- (D_true %*% beta_true) / L
    epsilon       <- matrix(rnorm(n * M, sd = 0.3), nrow = n)
    Y_mat         <- signal_scalar + signal_func + epsilon

    dat <- data.frame(age = age)
    dat$Y_mat  <- Y_mat
    dat$X_func <- X_func
    dat$sindex <- matrix(rep(sindex, n), nrow = n, byrow = TRUE)

## Fit the joint FPCA FoFR model

    fit_fofr_joint <- fofr_bayes(
      formula     = Y_mat ~ age + s(sindex, by = X_func, bs = "cr", k = 10),
      data        = dat,
      joint_FPCA  = c(TRUE),
      spline_type = "bs",
      spline_df   = 10,
      niter       = 2000,
      nwarmup     = 1000,
      nchain      = 3,
      ncores      = 3
    )

The joint FPCA contributes the additional Stan parameters `xi_1` (\(n \times J\)), `lambda_1` (\(J\)-vector), and `sigma_e_1` (scalar), and adds the FPCA log-likelihood and prior to the model block; the bivariate coefficient \(\hat\beta(s,t)\) is reconstructed from \(\mathbf{B}_r, \mathbf{B}_f\) exactly as in the no-joint FoFR case.

## Visualize the bivariate coefficient

    beta_est <- apply(fit_fofr_joint$bivar_func_coef[[1]], c(2, 3), mean)

    par(mfrow = c(1, 2), mar = c(4, 4, 2, 1))
    image(sindex, tindex, beta_true,
          xlab = "s (predictor domain)", ylab = "t (response domain)",
          main = expression("True " * beta(s,t)),
          col = hcl.colors(64, "Blue-Red 3"))
    image(sindex, tindex, beta_est,
          xlab = "s (predictor domain)", ylab = "t (response domain)",
          main = expression("Joint-FPCA " * hat(beta)(s,t)),
          col = hcl.colors(64, "Blue-Red 3"))

# Practical Recommendations

- When to use joint FPCA. Joint FPCA is most useful when the observed functional predictor \(W_i(s)\) contains substantial measurement noise relative to the signal in the regression, or when the curves are sparse / irregularly observed. For dense, low-noise functional data the difference is small.

- Number of FPC components \(J\). Controlled by the default variance-explained criterion in `refund::fpca.sc()`. A value such that cumulative variance explained exceeds 95 % is a reasonable default. A larger \(J\) increases the number of joint FPC score parameters (\(n \times J\)) and may slow down sampling.

- Convergence. Joint FPCA introduces \(n \times J\) extra latent parameters. Multiple chains and the standard `rstan` diagnostics (traceplots, \(\hat R\), bulk / tail ESS) are recommended:

      rstan::traceplot(fit_sofr_joint$stanfit, pars = c("sigma_e_1", "lambda_1"))
      print(fit_sofr_joint$stanfit, pars = c("sigma_e_1", "lambda_1"))

- Multiple functional predictors. The argument is a vector with one entry per `s(...)` functional term. Setting some entries to `TRUE` and others to `FALSE` is supported: each functional predictor is treated independently.

- Prior strength. The default `Inv-Gamma(0.001, 0.001)` priors on \(\lambda_j^{2}\) and \(\sigma_e^{2}\) are weakly informative. If the posterior of \(\xi_i\) is shrunk too hard toward \(\hat\xi_i\), consider loosening the prior by editing the Stan code (set `runStan = FALSE`, modify the `inv_gamma_lpdf(...)` lines, and pass the resulting program to `rstan::stan()` directly).

# References

- Jiang, Z., Crainiceanu, C., and Cui, E. (2025). Tutorial on Bayesian Functional Regression Using Stan. Statistics in Medicine, 44(20–22), e70265. (See Section 4 for the joint FPCA formulation that this vignette implements.)
- Crainiceanu, C. M., Goldsmith, J., Leroux, A., and Cui, E. (2024). Functional Data Analysis with R. CRC Press.
- Goldsmith, J., Bobb, J., Crainiceanu, C. M., Caffo, B., and Reich, D. (2011). Penalized Functional Regression. Journal of Computational and Graphical Statistics, 20(4), 830–851.
- Yao, F., Mueller, H.-G., and Wang, J.-L. (2005). Functional Data Analysis for Sparse Longitudinal Data. Journal of the American Statistical Association, 100(470), 577–590.
- Goldsmith, J., Greven, S., and Crainiceanu, C. M. (2013). Corrected Confidence Bands for Functional Data Using Principal Components. Biometrics, 69(1), 41–51.
- Carpenter, B., Gelman, A., Hoffman, M. D., et al. (2017). Stan: A Probabilistic Programming Language. Journal of Statistical Software, 76(1), 1–32.