# Missing Data and Partially Known Parameters

Bayesian inference supports a general approach to missing data in which any missing data item is represented as a parameter that is estimated in the posterior . If the missing data are not explicitly modeled, as in the predictors for most regression models, then the result is an improper prior on the parameter representing the missing predictor.

Mixing arrays of observed and missing data can be difficult to include in Stan, partly because it can be tricky to model discrete unknowns in Stan and partly because unlike some other statistical languages (for example, R and Bugs), Stan requires observed and unknown quantities to be defined in separate places in the model. Thus it can be necessary to include code in a Stan program to splice together observed and missing parts of a data structure. Examples are provided later in the chapter.

## Missing data

Stan treats variables declared in the data and transformed data blocks as known and the variables in the parameters block as unknown.

An example involving missing normal observations could be coded as follows.1

data {
int<lower=0> N_obs;
int<lower=0> N_mis;
array[N_obs] real y_obs;
}
parameters {
real mu;
real<lower=0> sigma;
array[N_mis] real y_mis;
}
model {
y_obs ~ normal(mu, sigma);
y_mis ~ normal(mu, sigma);
}

The number of observed and missing data points are coded as data with non-negative integer variables N_obs and N_mis. The observed data are provided as an array data variable y_obs. The missing data are coded as an array parameter, y_mis. The ordinary parameters being estimated, the location mu and scale sigma, are also coded as parameters. The model is vectorized on the observed and missing data; combining them in this case would be less efficient because the data observations would be promoted and have needless derivatives calculated.

## Partially known parameters

In some situations, such as when a multivariate probability function has partially observed outcomes or parameters, it will be necessary to create a vector mixing known (data) and unknown (parameter) values. This can be done in Stan by creating a vector or array in the transformed parameters block and assigning to it.

The following example involves a bivariate covariance matrix in which the variances are known, but the covariance is not.

data {
int<lower=0> N;
array[N] vector[2] y;
real<lower=0> var1;
real<lower=0> var2;
}
transformed data {
real<lower=0> max_cov = sqrt(var1 * var2);
real<upper=0> min_cov = -max_cov;
}
parameters {
vector[2] mu;
real<lower=min_cov, upper=max_cov> cov;
}
transformed parameters {
matrix[2, 2] Sigma;
Sigma[1, 1] = var1;
Sigma[1, 2] = cov;
Sigma[2, 1] = cov;
Sigma[2, 2] = var2;
}
model {
y ~ multi_normal(mu, Sigma);
}

The variances are defined as data in variables var1 and var2, whereas the covariance is defined as a parameter in variable cov. The $$2 \times 2$$ covariance matrix Sigma is defined as a transformed parameter, with the variances assigned to the two diagonal elements and the covariance to the two off-diagonal elements.

The constraint on the covariance declaration ensures that the resulting covariance matrix sigma is positive definite. The bound, plus or minus the square root of the product of the variances, is defined as transformed data so that it is only calculated once.

The vectorization of the multivariate normal is critical for efficiency here. The transformed parameter Sigma could be defined as a local variable within the model block if it does not need to be included in the sampler’s output.

## Sliced missing data

If the missing data are part of some larger data structure, then it can often be effectively reassembled using index arrays and slicing. Here’s an example for time-series data, where only some entries in the series are observed.

data {
int<lower=0> N_obs;
int<lower=0> N_mis;
array[N_obs] int<lower=1, upper=N_obs + N_mis> ii_obs;
array[N_mis] int<lower=1, upper=N_obs + N_mis> ii_mis;
array[N_obs] real y_obs;
}
transformed data {
int<lower=0> N = N_obs + N_mis;
}
parameters {
array[N_mis] real y_mis;
real<lower=0> sigma;
}
transformed parameters {
array[N] real y;
y[ii_obs] = y_obs;
y[ii_mis] = y_mis;
}
model {
sigma ~ gamma(1, 1);
y[1] ~ normal(0, 100);
y[2:N] ~ normal(y[1:(N - 1)], sigma);
}

The index arrays ii_obs and ii_mis contain the indexes into the final array y of the observed data (coded as a data vector y_obs) and the missing data (coded as a parameter vector y_mis). See the time series chapter for further discussion of time-series model and specifically the autoregression section for an explanation of the vectorization for y as well as an explanation of how to convert this example to a full AR(1) model. To ensure y[1] has a proper posterior in case it is missing, we have given it an explicit, albeit broad, prior.

Another potential application would be filling the columns of a data matrix of predictors for which some predictors are missing; matrix columns can be accessed as vectors and assigned the same way, as in

x[N_obs_2, 2] = x_obs_2;
x[N_mis_2, 2] = x_mis_2;

where the relevant variables are all hard coded with index 2 because Stan doesn’t support ragged arrays. These could all be packed into a single array with more fiddly indexing that slices out vectors from longer vectors (see the ragged data structures section for a general discussion of coding ragged data structures in Stan).

Rick Farouni, on the Stan users group, inquired as to how to build a Cholesky factor for a covariance matrix with a unit diagonal, as used in Bayesian factor analysis . This can be accomplished by declaring the below-diagonal elements as parameters, then filling the full matrix as a transformed parameter.

data {
int<lower=2> K;
}
transformed data {
int<lower=1> K_choose_2;
K_choose_2 = (K * (K - 1)) / 2;
}
parameters {
vector[K_choose_2] L_lower;
}
transformed parameters {
cholesky_factor_cov[K] L;
for (k in 1:K) {
L[k, k] = 1;
}
{
int i;
for (m in 2:K) {
for (n in 1:(m - 1)) {
L[m, n] = L_lower[i];
L[n, m] = 0;
i += 1;
}
}
}
}

It is most convenient to place a prior directly on L_lower. An alternative would be a prior for the full Cholesky factor L, because the transform from L_lower to L is just the identity and thus does not require a Jacobian adjustment (despite the warning from the parser, which is not smart enough to do the code analysis to infer that the transform is linear). It would not be at all convenient to place a prior on the full covariance matrix L * L', because that would require a Jacobian adjustment; the exact adjustment is detailed in the reference manual.

## Missing multivariate data

It’s often the case that one or more components of a multivariate outcome are missing.2

As an example, we’ll consider the bivariate distribution, which is easily marginalized. The coding here is brute force, representing both an array of vector observations y and a boolean array y_observed to indicate which values were observed (others can have dummy values in the input).

array[N] vector[2] y;
array[N, 2] int<lower=0, upper=1> y_observed;

If both components are observed, we model them using the full multi-normal, otherwise we model the marginal distribution of the component that is observed.

for (n in 1:N) {
if (y_observed[n, 1] && y_observed[n, 2]) {
y[n] ~ multi_normal(mu, Sigma);
} else if (y_observed[n, 1]) {
y[n, 1] ~ normal(mu[1], sqrt(Sigma[1, 1]));
} else if (y_observed[n, 2]) {
y[n, 2] ~ normal(mu[2], sqrt(Sigma[2, 2]));
}
}

It’s a bit more work, but much more efficient to vectorize these distribution statements. In transformed data, build up three vectors of indices, for the three cases above:

transformed data {
array[observed_12(y_observed)] int ns12;
array[observed_1(y_observed)] int ns1;
array[observed_2(y_observed)] int ns2;
}

You will need to write functions that pull out the count of observations in each of the three situations. This must be done with functions because the result needs to go in top-level block variable size declaration. Then the rest of transformed data just fills in the values using three counters.

int n12 = 1;
int n1 = 1;
int n2 = 1;
for (n in 1:N) {
if (y_observed[n, 1] && y_observed[n, 2]) {
ns12[n12] = n;
n12 += 1;
} else if (y_observed[n, 1]) {
ns1[n1] = n;
n1 += 1;
} else if (y_observed[n, 2]) {
ns2[n2] = n;
n2 += 1;
}
}

Then, in the model block, everything is vectorizable using those indexes constructed once in transformed data:

y[ns12] ~ multi_normal(mu, Sigma);
y[ns1] ~ normal(mu[1], sqrt(Sigma[1, 1]));
y[ns2] ~ normal(mu[2], sqrt(Sigma[2, 2]));

The result will be much more efficient than using latent variables for the missing data, but it requires the multivariate distribution to be marginalized analytically. It’d be more efficient still to precompute the three arrays in the transformed data block, though the efficiency improvement will be relatively minor compared to vectorizing the probability functions.

This approach can easily be generalized with some index fiddling to the general multivariate case. The trick is to pull out entries in the covariance matrix for the missing components. It can also be used in situations such as multivariate differential equation solutions where only one component is observed, as in a phase-space experiment recording only time and position of a pendulum (and not recording momentum).

1. A more meaningful estimation example would involve a regression of the observed and missing observations using predictors that were known for each and specified in the data block.↩︎