2.1 Setup and Estimands

We denote $i=1, 2,\ldots , N$ and $t=1, 2,\ldots , T$ as the unit and time for which and when the outcome of interest is observed. Although our method can accommodate imbalanced panels, we assume a balanced panel instead for notational convenience. We consider a binary treatment $w_{it}$ that, once it takes a value of 1, cannot reverse back to 0 (staggered adoption). Following Athey and Imbens (Reference Athey and Imbens2018), we define the timing of adoption for each unit i as a random variable $a_{i}$ that takes its values in $\mathbb {A}= \{1,2,\ldots , T, c\}$ , in which $a_{i} = c> T$ means that unit i falls in the residual category and does not get treated in the observed time window. We call unit i a treated unit if it adopts the treatment at any of the observed time periods ( $a_{i}=1, 2,\ldots , T$ ); we call it a control unit if it never adopts the treatment by period $T\ (a_{i} = c$ ). The number of pre-treatment periods for a treated unit is $T_{0,i} = a_{i} -1$ . Suppose there are $N_{co}$ control units and $N_{tr}$ treated units; $N_{co} + N_{tr} = N$ . In comparative case studies, $N_{tr}=1$ or a small integer.

For instance, in the German reunification example, $i=1,\ldots , 17$ , $T = 1, 2,\ldots , 44$ . West Germany’s $a_{i}$ is $31$ (the calendar year 1990); and $a_{i}>44$ for the other 16 OECD countries serving as controls. The EDR example uses the data of 47 states ( $i=1,\ldots , 47$ ) in 24 presidential election years from 1920 to 2012 ( $t=1,\ldots ,24)$ . Among the 47 states, 9 states adopted EDR before 2012, whose $a_{i} \in \{15, 20, 23, 24\}$ ; they are considered the treated units. The other 38 states did not adopt EDR by 2012 ( $a_{i}>24$ ) and are considered the control units.

Denote $\textbf{w}_{i} = (w_{i1},\ldots , w_{iT})^{\prime }$ as the treatment assignment vector for unit i. Staggered adoption implies that the adoption time $a_{i}$ uniquely determines vector $\textbf{w}_{i}$ : $\textbf{w}_{i}(a_{i})$ , in which ${w_{it}=0}$ if $t<a_{i}$ and ${w_{it}=1}$ if $t \geqslant a_{i}$ , for $t=1, 2,\ldots , T$ . We further define an $(N\times T)$ treatment assignment matrix, $ \textbf {W} =\{\textbf{w}_{1},\ldots , \textbf{w}_{N} \}$ ; similarly, $\textbf {W}$ is fully determined by $\mathcal {A}=\{a_{1},\ldots , a_{i},\ldots , a_{N} \}$ , the adoption time vector. Following Athey and Imbens (Reference Athey and Imbens2018), we make the following two assumptions to rule out cross-sectional spillover and the anticipation effect.

This assumption rules out cross-sectional spillover effects and significantly reduces the number of potential outcome trajectories. For each unit i, because now there are only $(T+1)$ possibilities for $\textbf{w}_{i}$ , there are $(T+1)$ potential outcome trajectories, denoted by $y_{it}(\textbf{w}_{i}), \ t = 1, 2,\ldots , T$ . In the German reunification example, this assumption rules out the possibility that German reunification affects economic growth in the other 16 countries, which may, in fact, be a strong assumption. In the EDR example, this assumption implies that the adoption of EDR laws in State B does not affect State A’s turnout with or without EDR laws. Because $\textbf{w}_{i}$ is fully determined by $a_{i}$ , we simply write $y_{it}(\textbf{w}_{i}(a_{i}))$ as $y_{it}(a_{i})$ .

Assumption 2 (No anticipation)

For all unit i, for all time periods before adoption $t < a_{i}$ :

$$ \begin{align*} y_{it}(a_{i}) = y_{it}(c), \mbox{ for } t < a_{i}, \forall i \end{align*} $$

in which $y_{it}(c)$ is the potential outcome under the “pure control” condition, that is, the treatment vector $\textbf{w}_{i}$ includes all zeros. This assumption says that the current untreated potential outcome does not depend on whether the unit gets the treatment in the future. The assumption is violated when anticipating that a unit will adopt the treatment in the future affects its outcome today. For example, if people anticipated German reunification to take place in 1990 and West Germany’s economy adjusted to that expectation before 1990, the assumption would be violated.

Estimands

Under Assumptions 1 and 2, for treated unit i whose adoption time $a_{i} \leqslant T$ , we define its treatment effect at $t \geqslant a_{i}$ as

$$ \begin{align*} \delta_{it} = y_{it}(a_{i}) - y_{it}(c), \mbox{ for } a_{i}\leqslant t \leqslant T. \end{align*} $$

In other words, we focus on the difference between the observed post-treatment outcome of treated unit i and the counterfactual outcome of the same unit that had never received the treatment by period $T\!$ . In the German reunification example, the causal effect of interest is the difference between the observed gross domestic product (GDP) per capita of West Germany in reunified Germany since 1990 and that of the counterfactual West Germany had it remained separated from East Germany.

Because $y_{it}(a_{i})$ of treated unit i is fully observed for $t\geqslant a_{i}$ , the Bayesian framework regards it as data. The counterfactual outcome $y_{it}(c)$ of treated unit i for $t\geqslant a_{i}$ , on the other hand, is an unknown quantity; we regard it as a random variable. We also define the sample average treatment effect on the treated (ATT) for units that have been under the treatment for a duration of p periods: $\delta _{p} = \frac {1}{N_{tr,p}} \sum _{i: T - p + 1 \leqslant a_{i}\leqslant T} \delta _{i, a_{i}+p-1}$ , where $N_{tr,p}$ is the number of treated units that have been treated for p periods in the sample.

Given that $y_{it}(a_{i})$ is observed in the post-treatment period, estimating $\delta _{it}$ is equivalent to constructing the counterfactual outcome $y_{it}(c)$ . Under Assumptions 1 and 2, we can denote $\textbf {Y}(\textbf {0})$ , a $(N\times T)$ matrix, as the potential outcome matrix under $\textbf{W}=\textbf {0}$ (i.e., $a_{i}=c, \forall i$ ). Given any realization of $\textbf{W}$ , we can partition the indices for $\textbf {Y}(\textbf {0})$ into two sets: $S_{0} \equiv \{(it)| w_{it} =0\}$ , with which $y_{it}(c)$ is observed; and $S_{1} \equiv \{(it)| w_{it} =1\}$ , with which $y_{it}(c)$ is missing. Additionally, $S=S_{0} \cup S_{1}$ . We denote the observed and missing parts of $\textbf {Y}(\textbf {0})$ as $\textbf {Y}(\textbf {0})^{obs}$ and $\textbf {Y}(\textbf {0})^{mis}$ , respectively, and $\textbf {Y}_{i}(\textbf {0})^{obs}$ and $\textbf {Y}_{i}(\textbf {0})^{mis}$ as the row vectors in $\textbf {Y}(\textbf {0})^{obs}$ and $\textbf {Y}(\textbf {0})^{mis}$ corresponding to unit i, respectively. $\textbf {X}_{it}$ is a $(p_{1} \times 1)$ vector of exogenous covariates. $\textbf {X}_{i} = (\textbf {X}_{i1}, \ldots , \textbf {X}_{iT})^{\prime }$ is a $(T\times p_{1})$ covariate matrix, and we define $\textbf {X} = \{\textbf {X}_{1}, \textbf {X}_{2}, \cdots , \textbf {X}_{N}\}$ .

2.2 The Assignment Mechanism

Rubin et al. (Reference Rubin, Wang, Yin, Zell, O’Hagen and West2010) lay down the fundamentals of Bayesian causal inference, which views “causal inference entirely as a missing data problem” (p. 685). In general, using the observed outcomes and covariates, as well as the assignment mechanism, we can stochastically impute counterfactuals from their posterior predictive distribution $\Pr (\textbf {Y}(\textbf{W})^{mis} |\textbf {X}, \textbf {Y}(\textbf{W})^{obs}, \textbf{W})$ . Since in this research the main causal quantity of interest is the (average) treatment effect on the treated, our primary goal is to predict counterfactuals $\textbf {Y}(\textbf {0})^{mis}$ , the untreated outcomes of the treated units. Because staggered adoption implies that $\textbf{W}$ is fully determined by $\mathcal {A}$ , the adoption time vector, we write the posterior predictive distribution of $\textbf {Y}(\textbf {0})^{mis}$ as

(1) $$ \begin{align} \Pr (\textbf{Y}(\textbf{0})^{mis}|\textbf{X}, \textbf{Y}(\textbf{0})^{obs}, \mathcal{A}) =& \frac{\Pr(\textbf{X}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs})\Pr(\mathcal{A}|\textbf{X}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs})}{\Pr(\textbf{X}, \textbf{Y}(\textbf{0})^{obs}, \mathcal{A})} \nonumber\\ \propto& \Pr(\textbf{X}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs}) \Pr(\mathcal{A}|\textbf{X}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs})\nonumber\\ \propto & \Pr(\textbf{X}, \textbf{Y}(\textbf{0})) \Pr(\mathcal{A}|\textbf{X}, \textbf{Y}(\textbf{0})). \end{align} $$

The Bayes rule gives the equality in Equation (1), and we obtain the proportionality by dropping the denominator as a normalizing constant since it contains no missing data. Hence, two component probabilities, the underlying “science” $\Pr (\textbf {X}, \textbf {Y}(\textbf {0}))$ and the treatment assignment mechanism $\Pr (\mathcal {A}|\textbf {X}, \textbf {Y}(\textbf {0}))$ , help predict the counterfactuals.

We assume that the treatment assignment is “individualistic” (Imbens and Rubin Reference Imbens and Rubin2015, 31), that is, the adoption time of unit i does not depend on the covariates or potential outcomes of other units or their time of adoption, given $\textbf{X}_{i}$ and $\textbf{Y}_{i}(\textbf {0})$ . We also require that each i has some nonzero chances of getting treated. The first part of this assumption is violated if policy diffusion takes place—for example, State A adopts EDR following State B’s policy shift—and such an emulation effect cannot be captured by a unit’s pre-treatment covariates and untreated potential outcomes. The positivity assumption means that all units in the sample have some probability of getting treated, thus justifying using control information to predict treated counterfactuals.

The fact that $\Pr (a_{i}|\textbf{X}_{i}, \textbf{Y}_{i}(\textbf {0})) =\Pr (a_{i}|\textbf{X}_{i}, \textbf{Y}_{i}(\textbf {0})^{obs}, \textbf{Y}_{i}(\textbf {0})^{mis})$ implies that the treatment assignment mechanism may be correlated with $\textbf{Y}_{i}(\textbf {0})^{mis}$ . To rule out potential confounding, one possibility is to impose further restrictions and make an ignorability assignment assumption $\Pr (a_{i}|\textbf{X}_{i}, \textbf{Y}_{i}(\textbf {0})^{obs})$ (Rubin Reference Rubin1978). However, in the more general setting of MNAR, this restriction is unlikely to be true. For instance, the timing of a state’s adopting an EDR law could be driven by legislators’ concern about future voter turnout in the absence of such laws. Therefore, we rely on the latent ignorability assumption to break the link between treatment assignment and control outcomes.

Assumption 4 (Latent ignorability)

Conditional on the observed pre-treatment covariates $\textbf{X}_{i}$ and a vector of latent variables $\textbf{U}_{i} = (u_{i1}, u_{i2}, \cdots , u_{iT})$ , the assignment mechanism is free from dependence on any missing or observed untreated outcomes for each unit i, that is,

(2) $$ \begin{align} \Pr(a_{i}|\textbf{X}_{i}, \textbf{Y}_{i}(\textbf{0}), \textbf{U}_{i}) = \Pr(a_{i}|\textbf{X}_{i}, \textbf{Y}_{i}(\textbf{0})^{mis}, \textbf{Y}_{i}(\textbf{0})^{obs}, \textbf{U}_{i}) = \Pr(a_{i}|\textbf{X}_{i}, \textbf{U}_{i}). \end{align} $$

Note that $\textbf{X}_{i}$ can include both time-varying and time-invariant pre-treatment covariates. $\textbf {U}_{i}$ captures both unit-level heterogeneity, such as unit fixed effects, and unit-specific time trends (e.g., $u_{it} = \gamma _{i} \cdot g(t)$ , in which $g(\cdot )$ is a function of time). We expect $\textbf {U}_{i}$ to be correlated with $\textbf{Y}_{i}(\textbf {0})$ ; in fact, we will extract it from $\textbf{Y}_{i}(\textbf {0})^{obs}$ . Therefore, once we condition on $\textbf{X}_{i}$ and $\textbf {U}_{i}$ , the entire time series of $\textbf{Y}_{i}(\textbf {0})$ is assumed to be independent of $a_{i}$ . Thus, we can understand Assumption 4 as an extension of the strict exogeneity assumption often assumed in fixed effects models. Like strict exogeneity, it rules out dynamic feedback from the past outcomes on current and future treatment assignments, conditional on $\textbf {U}_{i}$ .

The latent ignorability assumption is key to our approach. It is easy to see that the “parallel trends” assumption is implied by this assumption when $\textbf {U}_{i}$ is a unit-specific constant $u_{i1} = u_{i2} = \cdots = u_{iT} = u_{i}$ . What does this assumption mean substantively? In the EDR example, suppose there is an unmeasured downward trend in voter turnout in all states since the 1970s, driven by unknown socioeconomic forces. Its impact may be different across states due to differences in geography, demography, party organizations, ideological orientation, and so on. At the same time, the adoption of EDR in a state is correlated with how much such a trend would affect the state’s turnout in the absence of any policy changes. In this case, the “parallel time trends” assumption is invalid because $u_{it}$ is a time-varying confounder; however, we can correct for the biases if we can estimate, then condition on, the state-specific impact of this common trend. Hence, we further make a feasibility assumption.

This assumption is explicitly or implicitly made with the factor-augmented approach in the existing literature (Xu Reference Xu2017; Athey et al. Reference Athey, Bayati, Doudchenko, Imbens and Khosravi2018; Bai and Ng Reference Bai and Ng2020). It says that we can decompose the unit-specific time trends into multiple common trends with heterogeneous impacts. Assumption 5 is violated when $\textbf{U}_{1},\ldots , \textbf{U}_{n}$ have no common components; for example, when unit-specific time trends are idiosyncratic.

2.3 Posterior Predictive Inference

Under Assumption 4, we temporarily consider $\textbf{U}$ as part of the covariates and write $\textbf{X}$ and $\textbf{U}$ together as $\textbf{X}^{\prime }$ . Then we have the posterior predictive distribution of $\textbf {Y}(\textbf {0})^{mis}$ as

(3) $$ \begin{align} \Pr (\textbf{Y}(\textbf{0})^{mis}|\textbf{X}^{\prime}, \textbf{Y}(\textbf{0})^{obs}, \mathcal{A}) &\propto \Pr(\textbf{X}^{\prime}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs}) \Pr(\mathcal{A}|\textbf{X}^{\prime}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs})\nonumber\\ &\propto \Pr (\textbf{X}^{\prime}, \textbf{Y}(\textbf{0})^{mis}, \textbf{Y}(\textbf{0})^{obs}) \Pr(\mathcal{A}|\textbf{X}^{\prime}) \nonumber\\ &\propto \Pr(\textbf{X}^{\prime}, \textbf{Y}(\textbf{0})). \end{align} $$

The first line is simply to re-write Equation (1) by replacing $\textbf{X}$ with $\textbf{X}^{\prime }$ ; we reach the second step using the latent ignorability assumption; the last step drops the treatment assignment mechanism $\Pr (\mathcal {A}|\textbf{X}^{\prime })$ as a normalizing constant since it does not contain $\textbf {Y}(\textbf {0})^{mis}$ .

Equation (3) says that the latent ignorability assumption makes the treatment assignment mechanism ignorable in counterfactual prediction; as a result, we can ignore it as long as we condition on $\textbf{X}^{\prime }$ . In other words, under these assumptions, the task of developing the posterior predictive distributions of counterfactuals is reduced to model $\Pr (\textbf{X}^{\prime }, \textbf {Y}(\textbf {0})) $ . We need this trick because in comparative case studies, the number of treated units is small; as a result, we lack sufficient variation of the timing of adoption in data to model $\Pr (\mathcal {A}|\textbf{X}^{\prime })$ . To model the underlying “science,” we further make the assumption of exchangeability:

By de Finetti’s theorem (de Finetti Reference de Finetti, Kyburg and Smokler1963), $\{(\textbf{X}_{it}^{\prime }, y_{it}(c))\}$ can be written as i.i.d, given some parameters and their prior distributions. Note that $\Pr (\textbf{X}^{\prime }, \textbf {Y}(\textbf {0}))$ is equivalent to $\Pr (\{(\textbf{X}_{it}^{\prime }, y_{it}(c))\})$ , and we now can write the posterior predictive distribution of $\textbf {Y}(\textbf {0})^{mis}$ in Equation (3) as

(4) $$ \begin{align} \Pr (\textbf{Y}(\textbf{0})^{mis}|\textbf{X}^{\prime}, &\textbf{Y}(\textbf{0})^{obs}, \mathcal{A}) \propto \Pr(\{(\textbf{X}_{it}^{\prime}, y_{it}(c)) \}) \nonumber \\ \propto \int &\underbrace{\left(\prod_{it \in S_{1}} f(y_{it}(c)^{mis}|\textbf{X}_{it}, \boldsymbol{\theta}^{\prime}) \right)}_{{{\rm posterior predictive distribution}}} \underbrace{\left(\prod_{it \in S_{0}} f(y_{it}(c)^{obs}|\textbf{X}_{it}, \boldsymbol{\theta}^{\prime})\right)}_{{{\rm likelihood}}}\pi(\boldsymbol{\theta}) d \boldsymbol{\theta}, \end{align} $$

where $\boldsymbol {\theta }$ are the parameters that govern the DGP of $y_{it}(c)$ given $\textbf{X}^{\prime }_{it}$ , and $\boldsymbol {\theta }^{\prime }=(\boldsymbol {\theta }, \textbf{U})$ when we regard the latent covariates $\textbf{U}$ as parameters. Note that in Equation (4), the likelihood is based on observed outcomes while the posterior predictive distribution is for predicting the missing potential outcomes. The second proportionality is reached by de Finetti’s theorem and under the assumption that the set of parameters that govern the DGP of the covariates $\textbf{X}$ are independent of $\boldsymbol {\theta }$ . See Supplementary Material for a formal mathematical development of the posterior predictive distribution.

Recall that our objective is to impute the untreated potential outcomes for treated observations. If we assume that the DGPs of the outcomes in $S_{0}$ and $S_{1}$ follow the same functional form $f(\cdot )$ , we can build a parametric model and estimate parameters based on the likelihood and then predict $y_{it}(c)^{mis}$ for $i \leqslant N_{co}$ at $t\geqslant a_{i}$ using the posterior predictive distribution. If we can correctly estimate $\pi (\textbf{U}|\textbf{X}, \textbf {Y}(\textbf {0})^{obs})$ , the posterior distributions of $\textbf{U}$ , using a factor analysis, we can draw samples of treated counterfactuals $y_{it}(c)^{mis}$ from its posterior predictive distribution as in Equation (1) by integrating out the model parameters.

3.1 A Multilevel Model with Dynamic Factors

Assumption 7 (Functional form)

The untreated potential outcomes for unit $i =1,\ldots , N$ at $t=1,\ldots , T$ are specified as follows:

(5) $$ \begin{align} y_{it} (c)&= \textbf{X}_{it}^{\prime}\boldsymbol{\beta}_{it} + \boldsymbol{\gamma}_{i}^{\prime}\textbf{f}_{t} + \epsilon_{it}, \end{align} $$

(6) $$ \begin{align} \boldsymbol{\beta}_{it} &= \boldsymbol{\beta} + \boldsymbol{\alpha}_{i} +\boldsymbol{\xi}_{t}, \end{align} $$

(7) $$ \begin{align} \boldsymbol{\xi}_{t} &= \Phi_{\xi} \boldsymbol{\xi}_{t - 1} + \boldsymbol{e}_{t}, \ \textbf{f}_{t} = \Phi_{f} \textbf{f}_{t - 1} + \boldsymbol{\nu}_{t}. \end{align} $$

Equation (5) is the individual-level regression, in which $y_{it}(c)$ is explained by three components. The first one, $\textbf{X}_{it} \boldsymbol {\beta }_{it} $ , captures the relationships between observed covariates and the outcome. The double subscripts of $\boldsymbol {\beta }_{it}$ indicate that we allow these relationships to be heterogeneous across units and over time. Equation (6) decomposes $\boldsymbol {\beta }_{it}$ into three parts: $\boldsymbol {\beta }$ is the mean of $\boldsymbol {\beta }_{it}$ and shared by all observations, and $\boldsymbol {\alpha }_{i}$ and $\boldsymbol {\xi }_{t}$ are zero-mean unit- and time-specific “residuals” of $\boldsymbol {\beta }_{it}$ , respectively. The second component, $\boldsymbol {\gamma }_{i}^{\prime }\textbf{f}_{t} $ , is the multifactor term, in which $\textbf{f}_{t}$ and $\boldsymbol {\gamma }_{i}$ are $(r\times 1)$ vectors of factors and factor loadings, respectively. Consistent with Assumption 5, we use $\textbf{f}_{t}$ and $\boldsymbol {\gamma }_{i}$ to approximate $\textbf{U}_{i}$ . The last component, $\epsilon _{it}$ , represents $i.i.d.$ idiosyncratic errors. We further model the dynamics in $\boldsymbol {\xi }_{t}$ and $\textbf{f}_{t}$ by specifying autoregressive processes as shown in Equation (7). We assume both transition matrices $\Phi _{\xi }$ and $\Phi _{f}$ are diagonal: $\Phi _{\xi } = \textrm{Diag}(\phi _{\xi _{1}}, \ldots , \phi _{\xi _{p_{3}}}) $ and $\Phi _{f} = \textrm{Diag}(\phi _{f_{1}}, \ldots , \phi _{f_{r}})$ .Footnote ² Finally, we assume the individual- and group-level errors, $\epsilon _{it}$ , $\boldsymbol {e}_{t}$ , and $\boldsymbol {\nu }_{t}$ , to be i.i.d. normal.

The DM-LFM allows the slope coefficient of each covariate to vary by unit, time, both, or neither. To illustrate this flexibility, we rewrite the individual-level model in a reduced and matrix format a s

(8) $$ \begin{align} \boldsymbol{y}_{i}(c) = \boldsymbol{X}_{i}\boldsymbol{\beta} + \boldsymbol{Z}_{i}\boldsymbol{\alpha}_{i} + \boldsymbol{A}_{i} \boldsymbol{\xi} + \textbf{F}\boldsymbol{\gamma}_{i} + \boldsymbol{\epsilon}_{i}, \end{align} $$

in which $\textbf {F} = (\textbf{f}_{1}, \ldots , \textbf{f}_{T})^{\prime }$ is a $(T \times r)$ factor matrix. $\boldsymbol {Z}_{i}$ of dimension $(T\times p_{2})$ are covariates that have unit-specific slopes ( $\boldsymbol {\beta }+\boldsymbol {\alpha }_{i}$ ). $\boldsymbol {A}_{i}$ of dimension $(T \times p_{3})$ are covariates that have time-specific slopes ( $\boldsymbol {\beta }+\boldsymbol {\xi }_{t}$ ). Both $\boldsymbol {A}_{i}$ and $\boldsymbol {Z}_{i}$ are subsets of $\boldsymbol {X}_{i}$ . When the kth covariate is in $\boldsymbol {A}_{i} \cap \boldsymbol {Z}_{i}$ , it has a slope coefficient that varies across units and over time, that is, $\beta _{it}^{(k)} = \beta ^{(k)}+\alpha _{i}^{(k)}+\xi _{t}^{(k)}$ .  Accordingly, $\boldsymbol {\beta }$ is a $(p_{1}\times 1)$ coefficients vector, $\boldsymbol {\alpha }_{i} $ is a $(p_{2} \times 1)$ vector, $\boldsymbol {\xi } = (\boldsymbol {\xi }_{1}^{\prime }, \ldots , \boldsymbol {\xi }_{T}^{\prime })^{\prime }$ is a $(p_{3} \times 1)$ vector, and $p_{1} \geqslant p_{2}, p_{3}$ . Because $\boldsymbol {\alpha }_{i}, \boldsymbol {\xi }_{t}, \textbf{f}_{t}$ are centered at zero, the systemic part of the model is $\mathbb{E}[\textbf {y}_{i}(c)]=\textbf{X}_{i}\boldsymbol {\beta }$ . The rest of the components define the variance of the composite errors as $\boldsymbol {\Omega }_{\textbf {y}_{i}(c)-\boldsymbol {X}_{i}\boldsymbol {\beta }} = (\boldsymbol {Z}_{i}\boldsymbol {\alpha }_{i} + \boldsymbol {A}_{i} \boldsymbol {\xi } + \textbf {F}\boldsymbol {\gamma }_{i})^{\prime }(\boldsymbol {Z}_{i}\boldsymbol {\alpha }_{i} + \boldsymbol {A}_{i} \boldsymbol {\xi } + \textbf {F}\boldsymbol {\gamma }_{i})+ \sigma ^{2}_{\epsilon }\textbf {I}$ , which contains nonzero off-diagonal elements because of the components $\textbf {Z}_{i}$ and $\textbf {A}_{i}$ . Note that we allow $\boldsymbol {Z}_{i}\boldsymbol {\alpha }_{i}$ , $\boldsymbol {A}_{i}\boldsymbol {\xi }$ , and $\textbf {F}\boldsymbol {\gamma }_{i}$ to be arbitrarily correlated.

Figure 1 presents the model graphically, and the shrinkage parameters $\lambda $ ’s in the figure will be discussed later. Note that this outcome model governs $y_{it}(c)$ only. Without loss of generality, we can add $\delta _{it}w_{it} $ in the model in which $\delta _{it}$ is the causal effect for unit i at time t. This model is a dynamic and multilevel extension to several existing causal inference methods with TSCS data. For example, if we set $ \textbf {Z}_{i} = \textbf {A}_{i} = (1,1,\dots , 1)^{\prime }$ , and $ r = 0 $ , Equation (8) becomes a parametric linear DiD model with covariates and two-way fixed effects: $y_{it} = \delta _{it} w_{it} + \textbf{X}_{it}^{\prime } \boldsymbol {\beta } + \alpha _{i} + \xi _{t} + \epsilon _{it}$ . This is what Liu, Wang, and Xu (Reference Liu, Wang and Xu2020) call the fixed effects counterfactual model. When we put restrictions $\textbf {Z}_{it} = \varnothing $ and $\textbf{X}_{i} = \textbf {A}_{i}$ is time-invariant, our model is reduced to a factor model that justifies the SCM (Abadie, Diamond, and Hainmueller Reference Abadie, Diamond and Hainmueller2010): $y_{it} = \delta _{it} w_{it} + \xi _{t} + \textbf{X}_{i}^{\prime }\boldsymbol {\beta }_{t} + \boldsymbol {\gamma }_{i}^{\prime }\textbf{f}_{t} + \epsilon _{it}$ . Gsynth is also a special case of the model when we force the coefficients not to vary; that is, $\textbf {Z}_{it} = \varnothing $ and $\textbf {A}_{it} = \varnothing $ and $y_{it} = \delta _{it} w_{it} + \textbf{X}_{it}^{\prime }\boldsymbol {\beta } + \boldsymbol {\gamma }_{i}^{\prime }\textbf{f}_{t} + \epsilon _{it}$ (Xu Reference Xu2017).

Figure 1

A graphic representation of dynamic multilevel latent factor model (DM-LFM). The shaded nodes represent observed data, including untreated outcomes and covariates; the unshaded nodes represent “missing” data (treated counterfactuals) and parameters. Only one (treated) unit i is shown. $T_{0i} = a_{i}-1$ is the last period before the treatment starts to affect unit i. The focus of the graph is Period t. Covariates in periods other than t, as well as relationships between parameters and $y_{i1}(c)$ , $y_{iT_{0i}(c)}$ , and $y_{iT}(c)$ , are omitted for simplicity.

3.2 Bayesian Stochastic Model Specification Search

One advantage of the DM-LFM is that it is highly flexible. However, the large number of specification options poses a challenge to model selection. Bayesian stochastic model searching reduces the risks of model mis-specification and simultaneously incorporates model uncertainty. We use shrinkage priors to choose the number of latent factors and decide whether and how to include a covariate. Specifically, we adopt the Bayesian Lasso and Lasso-like hierarchical shrinkage methods based on recent research.Footnote ³ We apply the Bayesian Lasso shrinkage on $\boldsymbol {\beta }$ using the following hierarchical setting of mixture of a normal-exponential prior, $\beta _{k} | \tau _{\beta _{k}}^{2} \sim \mathcal {N}(0, \tau _{\beta _{k}}^{2}), \ \tau _{\beta _{k}}^{2} | \lambda _{\beta } \sim Exp(\frac {\lambda _{\beta }^{2}}{2}), \ \lambda _{\beta }^{2} \sim \mathcal {G}(a_{1}, a_{2}), \ k=1,\ldots , p_{1}$ . The tuning parameter $\lambda $ controls the sparsity and degree of shrinkage and can be understood as the Bayesian equivalent to the regulation penalty in a frequentist Lasso regression. Instead of fixing $\lambda $ at a single value, we take advantage of Bayesian hierarchical modeling and give it a Gamma distribution with hyper-parameters $a_{1}$ and $a_{2}$ .Footnote ⁴

To select the other components of the model, we impose shrinkage on $\boldsymbol {\alpha }_{i}$ , $\boldsymbol {\xi }_{t}$ , or $\boldsymbol {\gamma }_{i}$ to determine whether to include a $Z_{j}$ ( $\,j = 1, 2, \dots , p_{2}$ ), $A_{j}$ ( $\,j = 1, 2, \dots , p_{3}$ ), or $f_{j}$ ( $\,j = 1, 2, \dots , r$ ) in the model. We consider the Lasso-like hierarchical shrinkage approach with re-parameterization. Assume $\boldsymbol {\alpha }_{i}$ , $\boldsymbol {\gamma }_{i}$ , and $\boldsymbol {\xi }_{t}$ have diagonal variance–covariance matrices, $\textbf {H}_{0} = \textrm{Diag}(\omega _{\alpha _{1}}^{2}, \ldots , \omega _{\alpha _{p_{2}}}^{2}), \ \boldsymbol {\Gamma }_{0} = \textrm{Diag}(\omega _{\gamma _{1}}^{2}, \ldots , \omega _{\gamma _{r}}^{2}), \ \Sigma _{e} = \textrm{Diag}(\omega _{\xi _{1}}^{2}, \ldots , \omega _{\xi _{p_{3}}}^{2})$ , respectively. To have a shrinkage effect, we should allow the variance parameters $\omega ^{2}$ to have a positive probability to take the value zero. Therefore, we re-parameterize $\boldsymbol {\alpha }_{i}$ , $\boldsymbol {\xi }_{t}$ , and $\boldsymbol {\gamma }_{i}$ as $\boldsymbol {\alpha }_{i}=\boldsymbol {\omega }_{\boldsymbol {\alpha }} \cdot \tilde {\boldsymbol {\alpha }}_{i},\ \boldsymbol {\xi }_{t} =\boldsymbol {\omega }_{\boldsymbol {\xi }} \cdot \tilde {\boldsymbol {\xi }}_{t}, \ \boldsymbol {\gamma }_{i}=\boldsymbol {\omega }_{\boldsymbol {\gamma }} \cdot \tilde {\boldsymbol {\gamma }}_{i}$ , where $\boldsymbol {\omega }_{\alpha } = (\omega _{\alpha _{1}}, \ldots , \omega _{\alpha _{p_{2}}})^{\prime }$ , $\boldsymbol {\omega }_{\xi } = (\omega _{\xi _{1}}, \ldots , \omega _{\xi _{p_{3}}})^{\prime }$ , and $\boldsymbol {\omega }_{\gamma } = (\omega _{\gamma _{1}}, \ldots , \omega _{\gamma _{r}})^{\prime }$ are column vectors. After re-parameterization, the variances $\boldsymbol {\omega }^{2}$ appear in the model as coefficients $\boldsymbol {\omega }$ that can take values on the entire real line, and the new variance–covariance matrices become identity matrices.

Now we assign Lasso priors to each $\omega _{\alpha }$ , $\omega _{\xi }$ , and $\omega _{\gamma _{j}}$ to shrink varying parameters grouped by unit or time. Together with the shrinkage on $\boldsymbol {\beta }$ , the algorithm will decide de facto whether a certain covariate is included, whether its coefficient varies by time or across units, and how many latent factors are considered. Because the shrinkage priors do not have a point mass component at zero, parameters of less important covariates are not zeroed out completely. Instead, they stay in the model but with shrunk impacts and can be regarded as virtually excluded from the model. The posterior distribution of $\omega $ may be of different shapes. If it is clearly bimodal, it means that the associated parameter is included in the model; if it is close to unimodal and centered at zero, it means that the parameter is virtually excluded from the model; if, however, the posterior distribution of $\omega $ has three or more modes, it indicates that the data do not provide decisive information on whether the corresponding covariate or factor is sufficiently important—in some iterations, the parameter escapes the shrinkage while in others, it is trapped in a narrow neighborhood around zero. In each MCMC iteration, the algorithm samples a model consisting of the parameters that successfully escape the shrinkage, and posterior distributions of parameters generated by the stochastic search algorithm are based on a mixture of models in a continuous model space. In other words, this variable selection process is also a model-searching and model-averaging process.Footnote ⁵

The DF-LFM has many parameters, but most of them, listed in Table 1(b), are included as part of Bayesian parameter expansion for computational convenience or as hyper-parameters to adjust for parameter uncertainties. They do not appear in the likelihood or can be integrated out of the likelihood given the effective parameters. The effective parameters, including $\boldsymbol {\beta }$ and the variance parameters, are reported in Table 1(a). Their number is usually much smaller than the number of observations.

Table 1.

Total number of parameters.

The variance parameters for $\nu $ , e, $f_{t}$ do not appear in the table because we assume they have standard normal priors.

3.3 Implementing a DM-LFM

We develop an MCMC algorithm to estimate a DM-LFM. The core functions of the algorithm is written in C++. Due to space limitations, we present the details of choices of priors and the iterative steps of MCMC updating in Section A.2 in online Supplementary Material. Broadly speaking, implementing a DM-LFM takes the following three steps:

Step 1. Model searching and parameter estimation. We specify and estimate the DM-FLM model with Bayesian shrinkage to sample G draws (excluding draws in the burn-in stage) of the parameters from their posterior distributions, $\boldsymbol {\theta }^{(g)}_{it} \sim \pi (\boldsymbol {\theta }_{it}| \mathfrak {D})$ , where $\mathfrak {D}=\{(\textbf{X}_{it}, y_{it}(c)^{obs}): it \in S_{0}\}$ is the set of untreated observations. Because of Bayesian shrinkage, $\pi (\boldsymbol {\theta }_{it}| \mathfrak {D})$ is in effect a mixture of distributions.

Step 2. Prediction and integration. We conduct Bayesian prediction by generating draws of counterfactual $y_{it}(c)^{mis}$ for each treated unit at $a_{i} \leqslant t \leqslant T$ from its posterior predictive distribution: $f(y_{it}(c)^{mis}|\textbf{X}, \textbf {Y}(\textbf {0})^{obs}) \propto \int f(y_{it}(c)^{mis}|\textbf{X}_{it}, \boldsymbol {\theta }_{it}) \pi (\boldsymbol {\theta }_{it}|\mathfrak {D})d\boldsymbol {\theta }_{it}$ . Bayesian prediction is an empirical integration: a sample of the predicted counterfactual is generated by plugging each draw $\boldsymbol {\theta }^{(g)}_{it}$ from $\pi (\boldsymbol {\theta }_{it}| \mathfrak {D})$ into $f(y_{it}(c)^{mis}|\textbf{X}_{it}, \boldsymbol {\theta }_{it})$ to obtain $y^{(g)}_{it}(c)$ for $g=1,\ldots , G$ . The sample of counterfactuals is drawn from $f(y_{it}(c)^{mis}|\textbf{X}_{it},\mathfrak {D} )= f(y_{it}(c)^{mis}|\textbf{X}, \textbf {Y}^{obs}(\textbf {0}))$ without any unknown quantities.

Step 3. Inference and diagnostics. We make inference about the causal effect $\delta _{it}$ at $a_{i} \leqslant t \leqslant T$ for each treated unit i, by summarizing the empirical posterior distribution $\delta _{it}$ formed by $\delta ^{(g)}_{it} = y_{it}(a_{i}) - y^{(g)}_{it}(c), \ g=1,\ldots G $ . To summarize the results, we can obtain its posterior mean, variance, and the Bayesian 95% credibility interval. We can make inferences about other estimands, such as the ATT, by pooling the posterior draws of $\delta _{it}$ and summarizing their posterior distributions accordingly. We conduct Bayesian diagnostic tests on the convergence and mixing on main parameters’ posterior distributions and find that the MCMC algorithm converges fast and mixes well in our simulation and empirical studies.

4.1 A Simulated Example

We simulate a panel dataset of 50 units and 30 time periods based on the following DGP:

(9) $$ \begin{align} y_{it} = \delta_{it}w_{it} + \textbf{X}_{it}^{\prime}\boldsymbol{\beta}_{it} + {\boldsymbol{\gamma}}_{i}^{\prime} \textbf{f}_{t} + \epsilon_{it} = \delta_{it}w_{it} + \textbf{X}_{it}^{\prime}(\boldsymbol{\beta}+\boldsymbol{\alpha}_{i}+\boldsymbol{\xi}_{t}) + {\boldsymbol{\gamma}}_{i}^{\prime} \textbf{f}_{t} + \epsilon_{it} \end{align} $$

in which $w_{it}$ is the treatment indicator and $\delta _{it}$ is the treatment effect. $\textbf{X}_{it}$ is a vector of 10 covariates including an intercept and nine time-varying variables, but only the intercept and the first three covariates have nonzero, unit- and time-varying coefficients. We re-parameterize Equation (9) as $y_{it} = \delta _{it}w_{it} + \textbf{X}_{it}^{\prime }\boldsymbol {\beta } + \textbf{X}_{it}^{\prime } ( {\boldsymbol {\omega }}_{\boldsymbol {\alpha }} \cdot \tilde {\boldsymbol {\alpha }}_{i}) + \textbf{X}_{it}^{\prime } ( {\boldsymbol {\omega }}_{\textbf {x}} \cdot \tilde {\boldsymbol {\xi }}_{t}) + ({\boldsymbol {\omega }}_{\boldsymbol {\gamma }} \cdot \tilde {\boldsymbol {\gamma }}_{i})^{\prime } \textbf{f}_{t} + \epsilon _{it}$ such that $\tilde {\boldsymbol {\alpha }}_{i}$ , $\tilde {\boldsymbol {\xi }}_{t}$ , and $\tilde {\boldsymbol {\gamma }}_{i}$ all have univariances. Seven units receive the treatment starting from Period 21 and remain treated till Period 30. The remaining 23 units are never treated. The heterogeneous treatment effects are governed by $\delta _{it,t>20}= t -20+ \tau _{it}$ , in which $\tau _{it}\stackrel {i.i.d.}{\sim } N(0,1)$ . This means the expected value of the treatment effect gradually increases as the treatment duration grows, for example, from 1 in Period 21 to 10 in Period 30. The factor vector $\textbf{f}_{t}$ is two-dimensional and both factors follow an AR(1) process. The probability of getting treated is positively correlated with the sum of a unit’s factor loadings $\tilde {\boldsymbol {\gamma }}_{i}$ , which are also i.i.d. $N(0,1)$ . The selection on the factor loadings will cause biases in the causal estimates if a model does not account for the factor term or the covariates. We provide the details of the DGP in Supplementary Material.

Figure 2 shows the estimated ATT (posterior means) with their 95% credibility intervals using a Bayesian DiD model (a) and a DM-LFM model (b). The DiD model assumes $\boldsymbol {\omega }_{\boldsymbol {\alpha }}=\boldsymbol {\omega }_{\boldsymbol {\xi }}=\boldsymbol {\omega }_{\boldsymbol {\gamma }}=0$ ; in other words, the coefficients for the covariates are assumed to be constant, and no factors are included in the model. Figure 2(a) shows that with the DiD estimator, multiple estimates in the pre-treatment periods are away from zero and significant biases exist for the ATT estimates in the post-treatment periods. On the contrary, the DM-LFM performs significantly better: we do not observe any pre-trend and the ATT estimates are close to the true values, which are covered by the corresponding 95% credibility intervals.

Figure 2

Estimated average treatment effect on the treated (ATT): difference-in-differences (DiD) versus dynamic multilevel latent factor model (DM-LFM). The above figures show the ATT estimates and their 95% credibility intervals from the DiD and DM-LFM estimators, both implemented with Bayesian Markov Chain Monte Carlo (MCMC) algorithms. The red dashed lines represent the true ATT of the five treated units.

Figure 3(a) and Figure 3(b) show the invariant and time-varying components of the covariate coefficients, respectively. While $\boldsymbol {\beta }$ and $\boldsymbol {\xi }_{t}$ for the intercept and the first three covariates are clearly nonzero, the coefficients for the other six covariates are close to zero, which is consistent with the DGP. The algorithm also correctly selects the non-zero $\boldsymbol {\alpha }_{i}$ , the unit-varying components of coefficients (reported in Supplementary Material). Figure 3c shows the posterior distributions of $\boldsymbol {\omega }_{\gamma }$ , which measures the relative importance of each of the 10 factors subject to selection. It shows that two factors (Factors 1 and 2) have a $\omega $ with clear bimodal posteriors, and the others all have spike-shaped unimodal ones. This means that the model correctly identifies the number of factors to be two.

Figure 3

Posterior distributions of coefficients. (a) and (b) show the posterior mean and 95% credibility intervals of the invariant component and time-varying component of $\boldsymbol {\beta }_{it}$ , respectively. (c) shows the posterior distribution of each $\boldsymbol {\omega }_{\boldsymbol {\gamma }}$ , which captures the extent to which a factor influences the outcome.

This simulated example demonstrates that the DM-LFM performs well even when the sample size is relatively small and when researchers have limited knowledge about which covariates to put in, whether the covariates’ relationships with the outcome change over time or across units, or how many latent factors to include in the model.

4.2 Additional Monte Carlo Evidence

In addition, we conduct several sets of Monte Carlo exercises to study the properties of the proposed method and compare its relative performance against existing methods. Due to space limitations, we provide the details of these exercises in Supplementary Material and only briefly summarize two exercises and the major findings below.

In the first exercise, we study the role of each of the main components of a DM-LFM in estimation and investigate how the sample size affects the model performance. We simulate samples using the DGP as in Equation (9) while varying the sample size (both the total number of units N and the number of pre-treatment periods $T_{0}$ ). We estimate and compare the full DM-LFM model with its three simpler variants: a model with covariates but without factors, analogous to a DiD model including covariates with fixed coefficients; a model without covariates but with 10 latent factors, analogous to Gsynth without time-varying covariates; and a model with factors and covariates with fixed coefficients. Figure 4 shows the comparison when the number of units is relatively small ( $N=40$ ), and the full results are reported in Supplementary Material. In general, we find that DM-LFM outperforms the other three models in terms of bias, standard deviation, root mean squared errors (RMSE), and coverage. This exercise demonstrates that each of the key components of the model contributes to improving performance in causal effect estimation. The factor term seems to have the most impact, but covariates with varying coefficients also notably improve precision and coverage.

Figure 4

Comparison of model performance: root mean square errors (RMSE) and coverage. The above figures show the RMSE (a) and the coverage rate of 95% credibility intervals (b) of the average treatment effect on the treated (ATT) estimates using three sets of models: (1) models that do not include factors or covariates; (2) models that include 10 factors but no covariates; (3) models that include 10 factors and covariates with fixed coefficients; and (4) models that include 10 factors and covariates with varying coefficients.

The second exercise compares the performance of the DM-LFM with SCM and Gsynth in the case of a single treated unit. Tables A5–A7 in Supplementary Material show that, compared with the SCM, the DM-LFM has a much smaller RMSE. The DM-LFM outperforms Gsynth in the realistic scenario when the true number of factors is unknown, and when the number of factors is large and each of them produces relatively weak signals. Note that our Bayesian approach becomes significantly more computationally demanding as the sample size grows. For example, in one simulation in which $N= 50$ and $T_{0}=60$ , it takes about 30 seconds to run a DM-LFM on a 2019 6-core MacBook Pro while Gsynth takes less than a second.

5.1 Economic Impact of German Reunification

First, we build a DM-LFM incorporating all pre-treatment time-invariant covariates considered in ADH (Reference Abadie, Diamond and Hainmueller2015), including pre-treatment averages of trade openness, inflation rate, industry share, schooling, and investment rate. The initial model we consider is $y_{it}(c) = \textbf{X}_{i}(\boldsymbol {\beta } + \boldsymbol {\xi }_{t}) + \textbf{f}_{t}\boldsymbol {\gamma }_{i} + .pngilon _{it}$ , in which $\textbf{X}_{i}$ represents the time-invariant covariates. We do not include unit fixed effects or unit-varying coefficients to be consistent with the SCM, but time-varying coefficients are included. Figure A3 in Supplementary Material shows that the intercept has a strong time trend, but all the covariate coefficients are almost constant over time. Our result suggests four to six factors, with their $\omega $ ’s clearly having bimodal posteriors while several other factors exhibit mixed posteriors (Figure A4 in Supplementary Material).

We then produce an empirical posterior distribution of GDP per capita for counterfactual West Germany had German reunification not happened. To check the goodness-of-fit, we also calculate the model prediction of its GDP per capita in pre-treatment years. In Figure 5, we compare the counterfactual predictions of the SCM (a) with that of the DM-LFM (b), in which we shade the 95% Bayesian credibility intervals in gray. The dashed vertical line indicates the year 1989, 1 year before the adoption time. The two methods yield similar results: the GDP per capita of the counterfactual West Germany is higher than that of the actual West Germany during most of the post-reunification period except for the first few years after reunification.

Figure 5

Actual and estimated counterfactuals for West Germany. This figure shows the per capita gross domestic product (GDP) of actual West Germany and the posterior means (with 95% credibility intervals) of its untreated outcome from 1960 to 2003. The within-sample prediction is very precise and the interval is too narrow to be seen in the figure between 1960 and 1989.

Finally, we draw inferences about the treatment effects. Figure 6(a) and Figure 6(b) report the estimated effects of reunification on West Germany using the SCM and Bayesian DM-LFM, respectively. The corresponding 95% credibility intervals are added in (b). To lend further credibility to our causal estimates, we conduct a placebo test by setting 1987–1989, 3 years before reunification, as the placebo period. Figure 6(c) shows that the estimated effects at each time point during the placebo period are close to 0, buttressing our confidence in the identification assumptions.

Figure 6

Estimated effect of reunification on West Germany. This figure shows the estimated treatment effect of reunification on West Germany’s per capita gross domestic product (GDP) using the synthetic control method (SCM) (a) and dynamic multilevel latent factor model (DM-LFM) (b), as well as the result from a placebo test using DM-LFM in which the 1987–1989 period (before reunification) is taken as “treated” (c).

5.2 Election Day Registration and Voter Turnout

For this example, we specify a full DM-LFM model including the same time-varying covariates used in Xu (Reference Xu2017) (universal mail-in registration and motor voter registration) as well as 10 factors. Because there are only two covariates, we do not impose shrinkage on their $\boldsymbol {\beta }$ , but assign shrinkage priors to $\boldsymbol {\alpha }_{i}$ , $\boldsymbol {\xi }_{t}$ , and $\boldsymbol {\gamma }_{i}$ . Our result suggests that at least six factors affect outcome prediction (Figure A9 in Supplementary Material). In contrast, Gsynth only includes two factors using a leave-one-out cross-validation procedure. As for $\boldsymbol {\alpha }_{i}$ and $\boldsymbol {\xi }_{t}$ , the intercept varies in both time and space dimensions, but the varying parts of the slopes of the covariates are virtually shrunk to zero. We estimate the parameters using MCMC. Consistent with Xu (Reference Xu2017), we find that the two covariates do not explain much of the variation in turnout.

We then generate the posterior distributions of counterfactual outcomes for the nine treated states in their post-treatment years, based on which we estimate the effect of EDR on voter turnout. In Figure 7, we report the ATT for the same duration of adoption. To do so, we pool the posterior draws of the individual treatment effects of all treated states in $p^{th}$ year after adoption for $p = 1, 2, \dots , 6$ . Using the posterior distributions of $\hat {\delta }_{p}$ , we obtain their posterior means and Bayesian 95% credibility intervals.

Figure 7

The effect of election day registration (EDR) on voter turnout. This figure shows the estimated average treatment effect of EDR on voter turnout in the United States using Gsynth (Xu Reference Xu2017) (a) and dynamic multilevel latent factor model (DM-LFM) (b), respectively. On the x-axis, the positive integers indicate the duration of treatment while the pre-treatment years are labeled as nonpositive integers. Period 1 is the first presidential election year in which a state implements EDR. Because the treated states adopted EDR at different points in time, the number of treated units decreases as p increases.

Comparing the point and uncertainty estimates up to the sixth post-adoption presidential elections from Gsynth and the DM-LFM, the most notable difference between them is that the Bayesian 95% credibility intervals are considerably narrower than the 95% confidence intervals from Gsynth. We suspect that this is because our Bayesian approach has better predictive performance of individual counterfactuals than Gsynth as evidenced in Figure A11 in Supplementary Material, where we report the individual treatment effects on six treated states that have at least three post-treatment measures of the outcome using both Gsynth and the DM-LFM—our new method fits the trajectory of each treated unit in the pre-treatment period noticeably better than Gsynth.