# DIY modeling of a binary network outcome
Author: Peter Hoff
Date: 2024-02-20

#### Peter Hoff

#### 2024-02-20

The `amen` package provides wrapper functions `ame` and `ame_rep` for fitting a wide variety of AME models. However, wrapper functions can be opaque, and even wrapper functions with many options are limited by the energy and imagination of their authors. Therefore, some users may prefer to use some of the low-level `amen` functions to build their own custom Gibbs samplers for AME model fitting.

In this vignette, I show how to build such a sampler for a binary AME model. The example data we will use is the international relations data `IR90s` that comes with the `amen` package. We will model the binary indicator of conflict initiation between pairs of countries. Note that this dyadic variable is a count of the number of conflicts. For the purpose of this example, we dichotomize it:

    library(amen)

    Y<-1*(IR90s$dyadvars[,,"conflicts"] >0 )

    netplot(Y,plot.iso=FALSE,plotnames=TRUE)

Which pairs of countries have conflicts? Well, poor countries that are far away from each other don’t have conflicts. With this in mind, let’s fit a model of international conflict that has as predictors the GDPs of the initiator and target countries, as well as the geographic distance between the pair (both on a log scale):

    Xn<-log(IR90s$nodevars[,2])
    Xd<-log(IR90s$dyadvars[,,3]+1)

The binary probit AME model we will fit to `Y`, `Xn` and `Xd` is \[ \begin{align*} z_{i,j} & = \beta_0 + \beta_r \text{gdp}_i + \beta_c \text{gdp}_j + \beta_d \text{dist}_{i,j} + u_i^\top v_j + a_i +b_j + \epsilon_{i,j} \\ y_{i,j} & = 1( z_{i,j}> 0 ) \end{align*} \] where \(\text{Var}{ (\begin{smallmatrix} a_i \\ b_i \end{smallmatrix} ) } = \Sigma_{ab}\), \(\text{Var}{ (\begin{smallmatrix} u_i \\ v_i \end{smallmatrix} ) } = \Sigma_{uv}\), and \[ \text{Var}{ ( \begin{smallmatrix} \epsilon_{i,j} \\ \epsilon_{j,i}\end{smallmatrix}) } = \Sigma_\epsilon = \sigma^2 \begin{pmatrix} 1 & \rho \\ \rho & 1 \end{pmatrix}. \]

We will construct a Gibbs sampler to approximate the joint posterior distribution of all unknown quantities, which includes

- the global parameters \(\beta, \Sigma_{ab}, \Sigma_{uv}, \rho\);

- the latent nodal effects \(\{ (a_i, b_i, u_i, v_i): i=1,\ldots, n\}\);

- the unobserved latent dyadic variables \(z_{i,j}\).

Here we go. First note that the “linear regression” part of the model may be written as \(\beta^\top x_{i,j}\) where \(x_{i,j} = ( 1, \text{gdp}_i , \text{gdp}_j , \text{dist}_{i,j} )\). and \(\beta = ( \beta_0 , \beta_r, \beta_c \beta_d)\). We construct the “design array”, the \(n\times n \times p = 130 \times 130 \times 4\):

    X<-design_array(Xrow=Xn,Xcol=Xn,Xdyad=Xd,n=nrow(Y))

If you are not sure what this is doing, examine `X` yourself or study the following output:

    dim(X)

    ## [1] 130 130   4

    dimnames(X)[[3]]

    ## [1] "intercept" ".row"      ".col"      ".dyad"

    X[1:3,1:3,1]

    ##      [,1] [,2] [,3]
    ## [1,]    1    1    1
    ## [2,]    1    1    1
    ## [3,]    1    1    1

    X[1:3,1:3,2]

    ##          [,1]     [,2]     [,3]
    ## [1,] 2.959587 2.959587 2.959587
    ## [2,] 2.191654 2.191654 2.191654
    ## [3,] 4.894925 4.894925 4.894925

    X[1:3,1:3,3]

    ##          [,1]     [,2]     [,3]
    ## [1,] 2.959587 2.191654 4.894925
    ## [2,] 2.959587 2.191654 4.894925
    ## [3,] 2.959587 2.191654 4.894925

    X[1:3,1:3,4]

    ##          [,1]      [,2]      [,3]
    ## [1,] 0.000000 1.6733512 1.9257074
    ## [2,] 1.673351 0.0000000 0.9321641
    ## [3,] 1.925707 0.9321641 0.0000000

Next we need starting values for all (but one) of our unknown quantities. Because this is just a tutorial I will set these lazily:

    Sab<-diag(2)

    rho<-0

    R<-2

    Suv<-diag(2*R)

    U<-V<-matrix(0,nrow(Y),R)

    # just need Z[i,j]<0 if Y[i,j]=0, Z[i,j]>0 if Y[i,j]=1
    Z<-matrix(zscores(Y,ties.method="random"),nrow(Y),nrow(Y))
    Z<-Z-max(Z[!is.na(Y) & Y==0 ] )
    diag(Z)<-0

Finally, we specify some aspects of the prior distribution. The functions in the `amen` package use a \(g\)-prior for \(\beta\) by default (but this can be changed) and an arc-sine prior or uniform prior for \(\rho\). The priors for \(\Sigma_{ab}\) and \(\Sigma_{uv}\) are

- \(\Sigma_{ab}^{-1} \sim W( \Sigma_{ab0}^{-1}, \eta_0 )\);
- \(\Sigma_{uv}^{-1} \sim W( \Sigma_{uv0}^{-1} , \kappa_0)\).

Sparse network datasets may not provide much information with which to estimate \(\Sigma_{ab}\) and \(\Sigma_{uv}\). For such data, estimation stability can be improved by concentrating the scale of these covariance matrices near some empirical Bayes estimate, where the concentration is decreasing with the amount of information in the data. The following calculations provide one way to adaptively choose the priors for \(\Sigma_{ab}\) and \(\Sigma_{uv}\) based on the information in the data:

    ybar<-mean(Y,na.rm=TRUE) ; mu<-qnorm(ybar)
    E<- (Y - ybar)/dnorm(qnorm(ybar)) ; diag(E)<-0
    a<-rowMeans(E,na.rm=TRUE)  ; a[is.na(a)]<-0
    b<-colMeans(E,na.rm=TRUE)  ; b[is.na(b)]<-0

    vscale<-mean(diag(cov(cbind(a,b))))
    PHAT<-pnorm(mu+outer(a,b,"+"))
    vdfmlt<-.25/mean(PHAT*(1-PHAT))

    Sab0<-diag(2)*vscale
    eta0<-round(4*vdfmlt)

    Suv0<-diag(2*R)*vscale
    kappa0<-round((2*R+2)*vdfmlt)

Now all we need is a for-loop that iteratively updates the unknown quantities from their full conditional distributions:

    BETA<-NULL ; RHO<-NULL ; UV<-matrix(0,nrow(Y),nrow(Y))

    for(s in 1:10000)
    {
      # update beta, a and b
      tmp<-rbeta_ab_fc(Z,Sab,rho,X,offset=U%*%t(V))
      beta<-tmp$beta ; a<-tmp$a ; b<-tmp$b

      # update UV
      tmp<-rUV_fc(Z,U,V,Suv,rho,offset=Xbeta(X,beta) + outer(a,b,"+"))
      U<-tmp$U ; V<-tmp$V

      # update Suv
      Suv<-rSuv_fc(U,V,Suv0=Suv0,kappa0=kappa0)

      # update Sab
      Sab<-rSab_fc(a,b,Sab0=Sab0,eta0=eta0)

      # update rho
      rho<-rrho_mh(Z,rho,offset=Xbeta(X,beta)+outer(a,b,"+")+U%*%t(V))

      # update Z
      Z<-rZ_bin_fc(Z, Xbeta(X,beta)+outer(a,b,"+")+U%*%t(V), rho, Y)

      # periodically save some output
      if(s%%25==0 & s>500)
      {
        cat(round(100*s/10000),"% complete\n")
        BETA<-rbind(BETA,beta)
        RHO<-c(RHO,rho)
        UV<-UV+U%*%t(V)
        matplot(BETA,type="l")
      }
    }

The mixing is a bit slow, so you would probably want to run the Markov chain longer to obtain some final results.

Here is the “wrapper” version, fitting the same model:

    fit<-ame(Y,Xrow=Xn,Xcol=Xn,Xdyad=Xd,R=2,family="bin")

Now compare posterior density approximations:

    for(j in 1:4)
    {
      plot(density(BETA[,j] ),main="",xlab="")
      lines(density(fit$BETA[,j] ),col="green")
    }

The longer you run the Markov chains, the closer the two approximations should get to each other (and to the actual posterior distribution).