Method of the month: custom likelihoods with Stan

Once a month we discuss a particular research method that may be of interest to people working in health economics. We’ll consider widely used key methodologies, as well as more novel approaches. Our reviews are not designed to be comprehensive but provide an introduction to the method, its underlying principles, some applied examples, and where to find out more. If you’d like to write a post for this series, get in touch. This month’s method is custom likelihoods with Stan.

Principles

Regular readers of this blog will know that I am a fan of Bayesian methods. The exponential growth in personal computing power has opened up a whole new range of Bayesian models at home. WinBUGS and JAGS were the go-to pieces of software for estimating Bayesian models, both using Markov Chain Monte Carlo (MCMC) methods.  Theoretically, an MCMC chain will explore the posterior distribution. But MCMC has flaws. For example, if the target distribution has a high degree of curvature, such as many hierarchical models might exhibit, then MCMC chains can have trouble exploring it. To compensate, the chains stay in the ‘difficult’ bit of the space for longer before leaving to go elsewhere so its average oscillates around the true value. Asymptotically, these oscillations balance out, but in real, finite time, they ultimately lead to bias. And further, MCMC chains are very slow to converge to the target distribution, and for complex models can take a literal lifetime. An alternative, Hamiltonian Monte Carlo (HMC), provides a solution to these issues. Michael Betancourt’s introduction to HCM is great for anyone interested in the topic.

Stan is a ‘probabilistic programming language’ that implements HMC. A huge range of probability distributions are already implemented in the software, check out the manual for more information. And there is an R package, rstanarm, that estimates a number of standard models using normal R code that even means you can use these tools without learning the code. However, Stan may not have the necessary distributions for more complex econometric or statistical models. It used to be the case that you would have to build your own MCMC sampler – but given the problems with MCMC, this is now strongly advised against in lieu of HMC. Fortunately, we can implement our own probability density functions in Stan. So, if you can write down the (log) likelihood for your model, you can estimate it in Stan!

The aim of this post is to provide an example of implementing a custom probability function in Stan from the likelihood of our model. We will look at the nested logit model. These models have been widely used for multinomial choice problems. An area of interest among health economists is the choice of hospital provider. A basic multinomial choice model, such as a multinomial logit, requires an independence of irrelevant alternatives (IIA) assumption that says the odds of choosing one option over another is independent of any other alternative. For example, it would assume that the odds of me choosing the pharmacy in town over the hospital in town would be unaffected by a pharmacy opening on my own road. This is likely too strong. There are many ways to relax this assumption, the nested logit being one. The nested logit is useful when choices can be ‘nested’ in groups and assumes there is a correlation among choices with each group. For example, we can group health care providers into pharmacies, primary care clinics, hospitals, etc. such as this:

nested

Implementation

Econometric model

Firstly, we need a nesting structure for our choices, like that described above. We’ll consider a 2-level nesting structure, with branches and total choices, with Rt choices in each branch t. Like with most choice models we start from an additive random utility model, which is, for individual i=1,…,N, and with choice over branch and option:

U_{itr} = V_{itr} + \epsilon_{itr}

Then the chosen option is the one with the highest utility. The motivating feature of the nested logit is that the hierarchical structure allows us to factorise the joint probability of choosing branch and option r into a conditional and marginal model:

p_{itr} = p_{it} \times p_{ir|t}

Multinomial choice models arise when the errors are assumed to have a generalised extreme value (GEV) distribution, which gives use the multinomial logit model. We will model the deterministic part of the equation with branch-varying and option-varying variables:

V_{itr} = Z_{it}'\alpha + X_{itr}'\beta_t

Then the model can be written as:

p_{itr} = p_{it} \times p_{ir|t} = \frac{exp(Z_{it}'\alpha + \rho_t I_{it})}{\sum_{k \in T} exp(Z_{ik}'\alpha + \rho_k I_{ik})} \times \frac{exp(X_{itr}'\beta_t/\rho_t)}{\sum_{m \in R_t} exp( X_{itm}'\beta_t/\rho_t) }

where \rho_t is variously called a scale parameter, correlation parameter, etc. and defines the within branch correlation (arising from the GEV distribution). We also have the log-sum, which is also called the inclusive value:

I_{it} = log \left(  \sum_{m \in R_t} exp( X_{itm}'\beta_t/\rho_t)  \right).

Now we have our model setup, the log likelihood over all individuals is

\sum_{i=1}^N \sum_{k \in T} \sum_{m \in R_t} y_{itr} \left[ Z_{it}'\alpha + \rho_t I_{it} - log \left( \sum_{k \in T} exp(Z_{ik}'\alpha + \rho_k I_{ik}) \right) + X_{itr}'\beta_t/\rho_t - log \left(  \sum_{m \in R_t} exp( X_{itm}'\beta_t/\rho_t) \right) \right]

As a further note, for the model to be compatible with an ARUM specification, a number of conditions need to be satisfied. One of these is satisfied is 0<\rho_t \leq 1, so we will make that restriction. We have also only included alternative-varying variables, but we are often interested in individual varying variables and allowing parameters to vary over alternatives, which can be simply added to this specification, but we will leave them out for now to keep things “simple”. We will also use basic weakly informative priors and leave prior specification as a separate issue we won’t consider further:

\alpha \sim normal(0,5), \beta_t \sim normal(0,5), \rho_t \sim Uniform(0,1)

Software

DISCLAIMER: This code is unlikely to be the most efficient, nor can I guarantee it is 100% correct – use at your peril!

The following assumes a familiarity with Stan and R.

Stan programs are divided into blocks including data, parameters, and model. The functions block allows us to define custom (log) probability density functions. These take a form something like:

real xxx_lpdf(real y, ...){}

which says that the function outputs a real valued variable and take a real valued variable, y, as one of its arguments. The _lpdf suffix allows the function to act as a density function in the program (and equivalently _lpmf for log probability mass functions for discrete variables). Now we just have to convert the log likelihood above into a function. But first, let’s just consider what data we will be passing to the program:

  • N, the number of observations;
  • T, the number of branches;
  • P, the number of branch-varying variables;
  • Q, the number of choice-varying variables;
  • R, a T x 1 vector with the number of choices in each branch, from which we can also derive the total number of options as sum(R). We will call the total number of options Rk for now;
  • Y, a N x Rk vector, where Y[i,j] = 1 if individual i=1,…,N chose choice j=1,…,Rk;
  • Z, a N x T x P array of branch-varying variables;
  • X, a N x Rk x Q array of choice-varying variables.

And the parameters:

  • \rho , a T x 1 vector of correlation parameters;
  • \alpha , a P x 1 vector of branch-level covariates;
  • \beta , a P x T matrix of choice-varying covariates.

Now, to develop the code, we will specify the function for individual observations of Y, rather than the whole matrix, and then perform the sum over all the individuals in the model block. So we only need to feed in each individual’s observations into the function rather than the whole data set. The model is specified in blocks as follows (with all the data and parameter as arguments to the function):

functions{
 real nlogit_lpdf(real[] y, real[,] Z, real[,] X, int[] R, 
   vector alpha, matrix beta, vector tau){
//first define our additional local variables
 real lprob; //variable to hold log prob
 int count1; //keep track of which option in the loops
 int count2; //keep track of which option in the loops
 vector[size(R)] I_val; //inclusive values
 real denom; //sum denominator of marginal model
//for the variables appearing in sum loops, set them to zero
 lprob = 0;
 count1 = 0;
 count2 = 0;
 denom = 0;
 
 // determine the log-sum for each conditional model, p_ir|t, 
 //i.e. inclusive value
 for(k in 1:size(R)){
    I_val[k] = 0;
    for(m in 1:R[k]){
       count1 = count1 + 1;
       I_val[k] = I_val[k] + exp(to_row_vector(X[count1,])*
          beta[,k] /tau[k]);
       }
    I_val[k] = log(I_val[k]);
 }
 
 //determine the sum for the marginal model, p_it, denomininator
 for(k in 1:size(R)){
    denom = denom + exp(to_row_vector(Z[k,])*alpha + tau[k]*I_val[k]);
 }
 
 //put everything together in the log likelihood
 for(k in 1:size(R)){
    for(m in 1:R[k]){
       count2 = count2 + 1;
       lprob = lprob + y[count2]*(to_row_vector(Z[k,])*alpha + 
         tau[k]*I_val[k] - log(denom) + 
         to_row_vector(X[count2,])*beta[,k] - I_val[k]);
       }
    }
// return the log likelihood value
 return lprob;
 }
}
data{
 int N; //number of observations
 int T; //number of branches
 int R[T]; //number of options per branch
 int P; //dim of Z
 int Q; //dim of X
 real y[N,sum(R)]; //outcomes array
 real Z[N,T,P]; //branch-varying variables array
 real X[N,sum(R),Q]; //option-varying variables array
}
parameters{
 vector<lower=0, upper=1>[T] rho; //scale-parameters
 vector[P] alpha; //branch-varying parameters
 matrix[Q,T] beta; //option-varying parameters
}
model{
//specify priors
 for(p in 1:P) alpha[p] ~ normal(0,5); 
 for(q in 1:Q) for(t in 1:T) beta[q,t] ~ normal(0,5);

//loop over all observations with the data 
 for(i in 1:N){
    y[i] ~ nlogit(Z[i,,],X[i,,],R,alpha,beta,rho);
 }
}

Simulation model

To see whether our model is doing what we’re hoping it’s doing, we can run a simple test with simulated data. It may be useful to compare the result we get to those from other estimators; the nested logit is most frequently estimated using the FIML estimator. But, neither Stata nor R provide packages that estimate a model with branch-varying variables – another reason why we sometimes need to program our own models.

The code we’ll use to simulate the data is:

#### simulate 2-level nested logit data ###

N <- 300 #number of people
P <- 2 #number of branch variant variables
Q <- 2 #number of option variant variables
R <- c(2,2,2) #vector with number of options per branch
T <- length(R) #number of branches
Rk <- sum(R) #number of options

#simulate data

Z <- array(rnorm(N*T*P,0,0.5),dim = c(N,T,P))
X <- array(rnorm(N*Rk*Q,0,0.5), dim = c(N,Rk,Q))

#parameters
rho <- runif(3,0.5,1)
beta <- matrix(rnorm(T*Q,0,1),c(Q,T))
alpha <- rnorm(P,0,1)

#option models #change beta indexing as required
vals_opt <- cbind(exp(X[,1,]%*%beta[,1]/rho[1]),exp(X[,2,]%*%beta[,1]/rho[1]),exp(X[,3,]%*%beta[,2]/rho[2]),
 exp(X[,4,]%*%beta[,2]/rho[2]),exp(X[,5,]%*%beta[,3]/rho[3]),exp(X[,6,]%*%beta[,3]/rho[3]))

incl_val <- cbind(vals_opt[,1]+vals_opt[,2],vals_opt[,3]+vals_opt[,4],vals_opt[,5]+vals_opt[,6])

vals_branch <- cbind(exp(Z[,1,]%*%alpha + rho[1]*log(incl_val[,1])),
 exp(Z[,2,]%*%alpha + rho[2]*log(incl_val[,2])),
 exp(Z[,3,]%*%alpha + rho[3]*log(incl_val[,3])))

sum_branch <- rowSums(vals_branch)

probs <- cbind((vals_opt[,1]/incl_val[,1])*(vals_branch[,1]/sum_branch),
 (vals_opt[,2]/incl_val[,1])*(vals_branch[,1]/sum_branch),
 (vals_opt[,3]/incl_val[,2])*(vals_branch[,2]/sum_branch),
 (vals_opt[,4]/incl_val[,2])*(vals_branch[,2]/sum_branch),
 (vals_opt[,5]/incl_val[,3])*(vals_branch[,3]/sum_branch),
 (vals_opt[,6]/incl_val[,3])*(vals_branch[,3]/sum_branch))

Y = t(apply(probs, 1, rmultinom, n = 1, size = 1))

Then we’ll put the data into a list and run the Stan program with 500 iterations and 3 chains:

data <- list(
 y = Y,
 X = X,
 Z = Z,
 R = R,
 T = T,
 N = N,
 P = P,
 Q = Q
)

require(rstan)
rstan_options(auto_write = TRUE)
options(mc.cores = parallel::detectCores())

fit <- stan("C:/Users/Samuel/Dropbox/Code/nlogit.stan",
 data = data,
 chains = 3,
 iter = 500)

Which gives results (with 25t and 75th percentiles dropped to fit on screen):

> print(fit)
Inference for Stan model: nlogit.
3 chains, each with iter=500; warmup=250; thin=1; 
post-warmup draws per chain=250, total post-warmup draws=750.

             mean se_mean   sd    2.5%     50%     75%   97.5% n_eff Rhat
rho[1]       1.00    0.00 0.00    0.99    1.00    1.00    1.00   750  1
rho[2]       0.87    0.00 0.10    0.63    0.89    0.95    1.00   750  1
rho[3]       0.95    0.00 0.04    0.84    0.97    0.99    1.00   750  1
alpha[1]    -1.00    0.01 0.17   -1.38   -0.99   -0.88   -0.67   750  1
alpha[2]    -0.56    0.01 0.16   -0.87   -0.56   -0.45   -0.26   750  1
beta[1,1]   -3.65    0.01 0.32   -4.31   -3.65   -3.44   -3.05   750  1
beta[1,2]   -0.28    0.01 0.24   -0.74   -0.27   -0.12    0.15   750  1
beta[1,3]    0.99    0.01 0.25    0.48    0.98    1.15    1.52   750  1
beta[2,1]   -0.15    0.01 0.25   -0.62   -0.16    0.00    0.38   750  1
beta[2,2]    0.28    0.01 0.24   -0.16    0.28    0.44    0.75   750  1
beta[2,3]    0.58    0.01 0.24    0.13    0.58    0.75    1.07   750  1
lp__      -412.84    0.14 2.53 -418.56 -412.43 -411.05 -409.06   326  1

Samples were drawn using NUTS(diag_e) at Sun May 06 14:16:43 2018.
For each parameter, 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).

Which we can compare to the original parameters:

> beta
           [,1]       [,2]      [,3]
[1,] -3.9381389 -0.3476054 0.7191652
[2,] -0.1182806  0.2736159 0.5237470
> alpha
[1] -0.9654045 -0.6505002
> rho
[1] 0.9503473 0.9950653 0.5801372

You can see that the posterior means and quantiles of the distribution provide pretty good estimates of the original parameters. Convergence diagnostics such as Rhat and traceplots (not reproduced here) show good convergence of the chains. But, of course, this is not enough for us to rely on it completely – you would want to investigate further to ensure that the chains were actually exploring the posterior of interest.

Applications

I am not aware of any examples in health economics of using custom likelihoods in Stan. There are not even many examples of Bayesian nested logit models, one exception being a paper by Lahiri and Gao, who ‘analyse’ the nested logit using MCMC. But given the limitations of MCMC discussed above, one should prefer this implementation in the post rather than the MCMC samplers of that paper. It’s also a testament to computing advances and Stan that in 2001 an MCMC sampler and analysis could fill a paper in a decent econometrics journal and now we can knock one out for a blog post.

In terms of nested logit models in general in health economics, there are many examples going back 30 years (e.g. this article from 1987). More recent papers have preferred “mixed” or “random parameters” logit or probit specifications, which are much more flexible than the nested logit. We would advise these sorts of models for this reason. The nested logit was used as an illustrative example of estimating custom likelihoods for this post.

Credit

 

Method of the month: Semiparametric models with penalised splines

Once a month we discuss a particular research method that may be of interest to people working in health economics. We’ll consider widely used key methodologies, as well as more novel approaches. Our reviews are not designed to be comprehensive but provide an introduction to the method, its underlying principles, some applied examples, and where to find out more. If you’d like to write a post for this series, get in touch. This month’s method is semiparametric models with penalised splines.

Principles

A common assumption of regression models is that effects are linear and additive. However, nothing is ever really that simple. One might respond that all models are wrong, but some are useful, as George Box once said. And the linear, additive regression model has coefficients that can be interpreted as average treatment effects under the right assumptions. Sometimes though we are interested in conditional average treatment effects and how the impact of an intervention varies according to the value of some variable of interest. Often this relationship is not linear and we don’t know its functional form. Splines provide a way of estimating curves (or surfaces) of unknown functional form and are a widely used tool for semiparametric regression models. The term ‘spline’ was derived from the tool shipbuilders and drafters used to construct smooth edges: a bendable piece of material that when fixed at a number of points would relax into the desired shape.

Implementation

Our interest lies in estimating the unknown function m:

y_i = m(x_i) + e_i

A ‘spline’ in the mathematical sense is a function constructed piece-wise from polynomial functions. The places where the functions meet are known as knots and the spline has order equal to one more than the degree of the underlying polynomial terms. Basis-splines or B-splines are the typical starting point for spline functions. These are curves that are defined recursively as a sum of ‘basis functions’, which depend only on the polynomial degree and the knots. A spline function can be represented as a linear combination of B-splines, the parameters dictating this combination can be estimated using standard regression model estimation techniques. If we have N B-splines then our regression function can be estimated as:

y_i = \sum_{j=1}^N ( \alpha_j B_j(x_i) ) + e_i

by minimising \sum_{i=1}^N \{ y_i - \sum_{j=1}^N ( \alpha_j B_j(x_i) ) \} ^2. Where the B_j are the B-splines and the \alpha_j are coefficients to be estimated.

Useful technical explainers of splines and B-splines can be found here [PDF] and here [PDF].

One issue with fitting splines to data is that we run the risk of ‘overfitting’. Outliers might distort the curve we fit, damaging the external validity of conclusions we might make. To deal with this, we can enforce a certain level of smoothness using so-called penalty functions. The smoothness (or conversely the ‘roughness’) of a curve is often defined by the integral of the square of the second derivative of the curve function. Penalised-splines, or P-splines, were therefore proposed which added on this smoothness term multiplied by a smoothing parameter \lambda. In this case, we look to minimising:

\sum_{i=1}^N \{ y_i - \sum_{j=1}^N ( \alpha_j B_j(x_i) ) \}^2 + \lambda\int m''(x_i)^2 dx

to estimate our parameters. Many other different variations on this penalty have been proposed. This article provides a good explanation of P-splines.

An attractive type of spline has become the ‘low rank thin plate spline‘. This type of spline is defined by its penalty, which has a physical analogy with the resistance that a thin sheet of metal puts up when it is bent. This type of spline removes the problem associated with thin plate splines of having too many parameters to estimate by taking a ‘low rank’ approximation, and it is generally insensitive to the choice of knots, which other penalised spline regression models are not.

Crainiceanu and colleagues show how the low rank thin plate smooth splines can be represented as a generalised linear mixed model. In particular, our model can be represented as:

m(x_i) = \beta_0 + \beta_1x_i  + \sum_{k=1}^K u_k |x_i - \kappa_k|^3

where \kappa_k, k=1,...,K, are the knots. The parameters, \theta = (\beta_0,\beta_1,u_k)', can be estimated by minimising

\sum_{i=1}^N \{ y_i - m(x_i) \} ^2 + \frac{1}{\lambda} \theta ^T D \theta .

This is shown to give the mixed model

y_i = \beta_0 + \beta_1 + Z'b + u_i

where each random coefficient in the vector b is distributed as N(0,\sigma^2_b) and Z and D are given in the paper cited above.

As a final note, we have discussed splines in one dimension, but they can be extended to more dimensions. A two-dimensional spline can be generated by taking the tensor product of the two one dimensional spline functions. I leave this as an exercise for the reader.

Software

R

  • The package gamm4 provides the tools necessary for a frequentist analysis along the lines described in this post. It uses restricted maximum likelihood estimation with the package lme4 to estimate the parameters of the thin plate spline model.
  • A Bayesian version of this functionality is implemented in the package rstanarm, which uses gamm4 to produce the matrices for thin plate spline models and Stan for the estimation through the stan_gamm4 function.

If you wanted to implement these models for yourself from scratch, Crainiceanu and colleagues provide the R code to generate the matrices necessary to estimate the spline function:

n<-length(covariate)
X<-cbind(rep(1,n),covariate)
knots<-quantile(unique(covariate),
 seq(0,1,length=(num.knots+2))[-c(1,(num.knots+2))])
Z_K<-(abs(outer(covariate,knots,"-")))^3
OMEGA_all<-(abs(outer(knots,knots,"-")))^3
svd.OMEGA_all<-svd(OMEGA_all)
sqrt.OMEGA_all<-t(svd.OMEGA_all$v %*%
 (t(svd.OMEGA_all$u)*sqrt(svd.OMEGA_all$d)))
Z<-t(solve(sqrt.OMEGA_all,t(Z_K)))

Stata

I will temper this advice by cautioning that I have never estimated a spline-based semi-parametric model in Stata, so what follows may be hopelessly incorrect. The only implementation of penalised splines in Stata is the package and associated function psplineHowever, I cannot find any information about the penalty function used, so I would advise some caution when implementing. An alternative is to program the model yourself, through conversion of the above R code in Mata to generate the matrix Z and then the parameters could be estimated with xtmixed. 

Applications

Applications of these semi-parametric models in the world of health economics have tended to appear more in technical or statistical journals than health economics journals or economics more generally. For example, recent examples include Li et al who use penalised splines to estimate the relationship between disease duration and health care costs. Wunder and co look at how reported well-being varies over the course of the lifespan. And finally, we have Stollenwerk and colleagues who use splines to estimate flexible predictive models for cost-of-illness studies with ‘big data’.

Credit

 

Method of the month: Synthetic control

Once a month we discuss a particular research method that may be of interest to people working in health economics. We’ll consider widely used key methodologies, as well as more novel approaches. Our reviews are not designed to be comprehensive but provide an introduction to the method, its underlying principles, some applied examples, and where to find out more. If you’d like to write a post for this series, get in touch. This month’s method is synthetic control.

Principles

Health researchers are often interested in estimating the effect of a policy of change at the aggregate level. This might include a change in admissions policies at a particular hospital, or a new public health policy applied to a state or city. A common approach to inference in these settings is difference in differences (DiD) methods. Pre- and post-intervention outcomes in a treated unit are compared with outcomes in the same periods for a control unit. The aim is to estimate a counterfactual outcome for the treated unit in the post-intervention period. To do this, DiD assumes that the trend over time in the outcome is the same for both treated and control units.

It is often the case in practice that we have multiple possible control units and multiple time periods of data. To predict the post-intervention counterfactual outcomes, we can note that there are three sources of information: i) the outcomes in the treated unit prior to the intervention, ii) the behaviour of other time series predictive of that in the treated unit, including outcomes in similar but untreated units and exogenous predictors, and iii) prior knowledge of the effect of the intervention. The latter of these only really comes into play in Bayesian set-ups of this method. With longitudinal data we could just throw all this into a regression model and estimate the parameters. However, generally, this doesn’t allow for unobserved confounders to vary over time. The synthetic control method does.

Implementation

Abadie, Diamond, and Haimueller motivate the synthetic control method using the following model:

y_{it} = \delta_t + \theta_t Z_i + \lambda_t \mu_i + \epsilon_{it}

where y_{it} is the outcome for unit i at time t, \delta_t are common time effects, Z_i are observed covariates with time-varying parameters \theta_t, \lambda_t are unobserved common factors with \mu_i as unobserved factor loadings, and \epsilon_{it} is an error term. Abadie et al show in this paper that one can derive a set of weights for the outcomes of control units that can be used to estimate the post-intervention counterfactual outcomes in the treated unit. The weights are estimated as those that would minimise the distance between the outcome and covariates in the treated unit and the weighted outcomes and covariates in the control units. Kreif et al (2016) extended this idea to multiple treated units.

Inference is difficult in this framework. So to produce confidence intervals, ‘placebo’ methods are proposed. The essence of this is to re-estimate the models, but using a non-intervention point in time as the intervention date to determine the frequency with which differences of a given order of magnitude are observed.

Brodersen et al take a different approach to motivating these models. They begin with a structural time-series model, which is a form of state-space model:

y_t = Z'_t \alpha_t + \epsilon_t

\alpha_{t+1} = T_t \alpha_t + R_t \eta_t

where in this case, y_t is the outcome at time t, \alpha_t is the state vector and Z_t is an output vector with \epsilon_t as an error term. The second equation is the state equation that governs the evolution of the state vector over time where T_t is a transition matrix, R_t is a diffusion matrix, and \eta_t is the system error.

From this setup, Brodersen et al expand the model to allow for control time series (e.g. Z_t = X'_t \beta), local linear time trends, seasonal components, and allowing for dynamic effects of covariates. In this sense the model is perhaps more flexible than that of Abadie et al. Not all of the large number of covariates may be necessary, so they propose a ‘slab and spike’ prior, which combines a point mass at zero with a weakly informative distribution over the non-zero values. This lets the data select the coefficients, as it were.

Inference in this framework is simpler than above. The posterior predictive distribution can be ‘simply’ estimated for the counterfactual time series to give posterior probabilities of differences of various magnitudes.

Software

Stata

  • Synth Implements the method of Abadie et al.

R

  • Synth Implements the method of Abadie et al.
  • CausalImpact Implements the method of Brodersen et al.

Applications

Kreif et al (2016) estimate the effect of pay for performance schemes in hospitals in England and compare the synthetic control method to DiD. Pieters et al (2016) estimate the effects of democratic reform on under-five mortality. We previously covered this paper in a journal round-up and a subsequent post, for which we also used the Brodersen et al method described above. We recently featured a paper by Lépine et al (2017) in a discussion of user fees. The synthetic control method was used to estimate the impact that the removal of user fees had in various districts of Zambia on use of health care.

Credit