In the vignette on the model definition, we pointed out that the
probit model can capture choice behavior heterogeneity by imposing a
mixing distribution on the coefficient vector. The implementation in
{RprobitB} is explained in this vignette^{1}.

The {mlogit} package (Croissant
2020) contains the data set `Electricity`

, in
which residential electricity customers were asked to decide between
four hypothetical electricity suppliers. The suppliers differed in 6
characteristics:

- their fixed price
`pf`

per kWh, - their contract length
`cf`

, - a boolean
`loc`

, indicating whether the supplier is a local company, - a boolean
`wk`

, indicating whether the supplier is a well known company, - a boolean
`tod`

, indicating whether the supplier offers a time-of-day electricity price (which is higher during the day and lower during the night), and - a boolean
`seas`

, indicating whether the supplier’s price is seasonal dependent.

This constitutes a choice situation where choice behavior
heterogeneity is expected: some customers might prefer a time-of-day
electricity price (because they may be not at home during the day),
while others can have the opposite preference. Ideally these differences
in preferences should be modeled using characteristics of the deciders.
In many cases (as in this data set) we do not have adequate information.
Instead these differences in taste can be captured by means of a mixing
distribution for the `tod`

coefficient. This corresponds to
the assumption of a random coefficient from the underlying mixing
distribution to be drawn for each decider. We can use the estimated
mixing distribution to determine for example the share of deciders that
have a positive versus negative preference towards time-of-day
electricity prices.

Additionally, we expect correlations between the random coefficients
to certain covariates, for example a positive correlation between the
influence of `loc`

and `wk`

: deciders that prefer
local suppliers might also prefer well known companies due to
recommendations and past experiences, although they might be more
expensive than unknown suppliers. The fitted multivariate normal
distribution will reveal these correlations.

The following lines prepare the `Electricity`

data set for
estimation. We use the convenience function `as_cov_names()`

that relabels the data columns for alternative specific covariates into
the required format
“`<covariate>_<alternative>`

”, compare to the
vignette on choice data. Via the
`re = c("cl","loc","wk","tod","seas")`

argument, we specify
that we want to model random effects for all but the price coefficient,
which we will fix to `-1`

to interpret the other estimates as
monetary values.

```
data("Electricity", package = "mlogit")
<- as_cov_names(Electricity, c("pf","cl","loc","wk","tod","seas"), 1:4)
Electricity <- prepare_data(
data form = choice ~ pf + cl + loc + wk + tod + seas | 0,
choice_data = Electricity,
re = c("cl","loc","wk","tod","seas")
)<- fit_model(data, R = 1000, scale = "pf := -1") model_elec
```

Calling the `coef()`

method on the estimated model also
returns the estimated (marginal) variances of the mixing distribution
besides the average mean effects:

```
coef(model_elec)
#> Estimate (sd) Variance (sd)
#> 1 pf -1.00 (0.00) NA (NA)
#> 2 cl -0.27 (0.04) 0.33 (0.04)
#> 3 loc 2.89 (0.27) 7.82 (1.60)
#> 4 wk 2.14 (0.22) 4.15 (0.97)
#> 5 tod -9.91 (0.25) 13.01 (2.29)
#> 6 seas -9.96 (0.21) 6.73 (1.45)
```

By the sign of the estimates we can for example deduce, that the
existence of the time-of-day electricity price `tod`

in the
contract has a negative effect. However, the deciders are very
heterogeneous here, because the estimated variance of this coefficient
is large. The same holds for the contract length `cl`

. In
particular, the estimated share of the population that prefers to have a
longer contract length equals:

```
<- coef(model_elec)["cl","mean"]
cl_mu <- sqrt(coef(model_elec)["cl","var"])
cl_sd pnorm(cl_mu / cl_sd)
#> [1] 0.3185047
```

The correlation between the covariates can be accessed as follows:^{2}

```
cov_mix(model_elec, cor = TRUE)
#> cl loc wk tod seas
#> cl 1.00000000 0.09161238 0.058215654 -0.05419413 -0.100685302
#> loc 0.09161238 1.00000000 0.792512872 0.13606295 0.027063294
#> wk 0.05821565 0.79251287 1.000000000 0.13506495 0.003499879
#> tod -0.05419413 0.13606295 0.135064950 1.00000000 0.544640146
#> seas -0.10068530 0.02706329 0.003499879 0.54464015 1.000000000
```

Here, we see the confirmation of our initial assumption about a high
correlation between `loc`

and `wk`

. The pairwise
mixing distributions can be visualized via calling the
`plot()`

method with the additional argument
`type = mixture`

:

`plot(model_elec, type = "mixture")`

More generally, {RprobitB} allows to specify a Gaussian mixture as the mixing distribution. In particular,

\[ \beta \sim \sum_{c=1}^C \text{MVN} (b_c,\Omega_c).\] This specification allows for a) a better approximation of the true underlying mixing distribution and b) a preference based classification of the deciders.

To estimate a latent mixture, specify a named list
`latent_classes`

with the following arguments and submit it
to the estimation routine `fit_model`

:

`C`

, the fixed number (greater or equal 1) of latent classes, which is set to 1 per default,^{3}`weight_update`

, a boolean, set to`TRUE`

for a weight-based update of the latent classes, see below,`dp_update`

, a boolean, set to`TRUE`

for a Dirichlet process-based update of the latent classes, see below,`Cmax`

, the maximum number of latent classes, set to`10`

per default.

The following weight-based updating scheme is analogue to Bauer, Büscher, and Batram (2019) and executed within the burn-in period:

We remove class \(c\), if \(s_c<\varepsilon_{\text{min}}\), i.e. if the class weight \(s_c\) drops below some threshold \(\varepsilon_{\text{min}}\). This case indicates that class \(c\) has a negligible impact on the mixing distribution.

We split class \(c\) into two classes \(c_1\) and \(c_2\), if \(s_c>\varepsilon_\text{max}\). This case indicates that class \(c\) has a high influence on the mixing distribution whose approximation can potentially be improved by increasing the resolution in directions of high variance. Therefore, the class means \(b_{c_1}\) and \(b_{c_2}\) of the new classes \(c_1\) and \(c_2\) are shifted in opposite directions from the class mean \(b_c\) of the old class \(c\) in the direction of the highest variance.

We join two classes \(c_1\) and \(c_2\) to one class \(c\), if \(\lVert b_{c_1} - b_{c_2} \rVert<\varepsilon_{\text{distmin}}\), i.e. if the euclidean distance between the class means \(b_{c_1}\) and \(b_{c_2}\) drops below some threshold \(\varepsilon_{\text{distmin}}\). This case indicates location redundancy which should be repealed. The parameters of \(c\) are assigned by adding the values of \(s\) from \(c_1\) and \(c_2\) and averaging the values for \(b\) and \(\Omega\).

These rules contain choices on the values for \(\varepsilon_{\text{min}}\), \(\varepsilon_{\text{max}}\) and \(\varepsilon_{\text{distmin}}\). The adequate value for \(\varepsilon_{\text{distmin}}\) depends on the scale of the parameters. Per default, {RprobitB} sets

`epsmin = 0.01`

,`epsmax = 0.99`

, and`distmin = 0.1`

.

These values can be adapted through the `latent_class`

list.

As an alternative to the weight-based updating scheme to determine
the correct number \(C\) of latent
classes, {RprobitB} implemented the Dirichlet process.^{4} The Dirichlet Process
is a Bayesian nonparametric method, where nonparametric means that the
number of model parameters can theoretically grow to infinity. The
method allows to add more mixture components to the mixing distribution
if needed for a better approximation, see Neal
(2000) for a
documentation of the general case. The literature offers many
representations of the method, including the Chinese Restaurant Process
(Aldous 1985), the stick-braking methapor
(Sethuraman 1994), and the Polya Urn
model (Blackwell and MacQueen 1973).

In our case, we face the situation to find a distribution \(g\) that explains the decider-specific coefficients \((\beta_n)_{n = 1,\dots,N}\), where \(g\) is supposed to be a mixture of an unknown number \(C\) of Gaussian densities, i.e. \(g = \sum_{c = 1,\dots,C} s_c \text{MVN}(b_c, \Omega_c)\).

Let \(z_n \in \{1,\dots,C\}\) denote the class membership of \(\beta_n\). A priori, the mixture weights \((s_c)_c\) are given a Dirichlet prior with concentration parameter \(\delta/C\), i.e. \((s_c)_c \mid \delta \sim \text{D}_C(\delta/C,\dots,\delta/C)\). Rasmussen (2000) shows that

\[ \Pr((z_n)_n\mid \delta) = \frac{\Gamma(\delta)}{\Gamma(N+\delta)} \prod_{c=1}^C \frac{\Gamma(m_c + \delta/C)}{\Gamma(\delta/C)}, \] where \(\Gamma(\cdot)\) denotes the gamma function and \(m_c = \#\{n:z_n = c\}\) the number of elements that are currently allocated to class \(c\). Crucially, the last equation is independent of the class weights \((s_c)_c\), yet it still depends on the finite number \(C\) of latent classes. However, Li, Schofield, and Gönen (2019) shows that

\[ \Pr(z_n = c \mid z_{-n}, \delta) = \frac{m_{c,-n} + \delta/C}{N-1+\delta},\] where the notation \(-n\) means excluding the \(n\)th element. We can let \(C\) approach infinity to derive:

\[ \Pr(z_n = c \mid z_{-n}, \delta) \to \frac{m_{c,-n}}{N-1+\delta}. \]

Note that the allocation probabilities do not sum to 1, instead

\[ \sum_{c = 1}^C \frac{m_{c,-n}}{N-1+\delta} = \frac{N-1}{N-1+\delta}. \]

The difference to 1 equals

\[ \Pr(z_n \neq z_m ~ \forall ~ m \neq n \mid z_{-n}, \delta) = \frac{\delta}{N-1+\delta} \]

and constitutes the probability that a new cluster for observation \(n\) is created. Neal (2000) points out that this probability is proportional to the prior parameter \(\delta\): A greater value for \(\delta\) encourages the creation of new clusters, a smaller value for \(\delta\) increases the probability of an allocation to an already existing class.

In summary, the Dirichlet process updates the allocation of each \(\beta\) coefficient vector one at a time, dependent on the other allocations. The number of clusters can theoretically rise to infinity, however, as we delete unoccupied clusters, \(C\) is bounded by \(N\). As a final step after the allocation update, we update the class means \(b_c\) and covariance matrices \(\Omega_c\) by means of their posterior predictive distribution. The mean and covariance matrix for a new generated cluster is drawn from the prior predictive distribution. The corresponding formulas are given in Li, Schofield, and Gönen (2019).

The Dirichlet process directly integrates into our existing Gibbs
sampler. Given \(\beta\) values, it
updated the class means \(b_c\) and
class covariance matrices \(\Omega_c\).
The Dirichlet process updating scheme is implemented in the function
`update_classes_dp()`

. In the following, we give a small
example in the bivariate case `P_r = 2`

. We sample true class
means `b_true`

and class covariance matrices
`Omega_true`

for `C_true = 3`

true latent classes.
The true (unbalanced) class sizes are given by the vector
`N`

, and `z_true`

denotes the true
allocations.

```
set.seed(1)
<- 2
P_r <- 3
C_true <- c(100,70,30)
N <- matrix(replicate(C_true, rnorm(P_r)), nrow = P_r, ncol = C_true))
(b_true #> [,1] [,2] [,3]
#> [1,] -0.6264538 -0.8356286 0.3295078
#> [2,] 0.1836433 1.5952808 -0.8204684
<- matrix(replicate(C_true, rwishart(P_r + 1, 0.1*diag(P_r))$W, simplify = TRUE),
(Omega_true nrow = P_r*P_r, ncol = C_true))
#> [,1] [,2] [,3]
#> [1,] 0.3093652 0.14358543 0.2734617
#> [2,] 0.1012729 -0.07444148 -0.1474941
#> [3,] 0.1012729 -0.07444148 -0.1474941
#> [4,] 0.2648235 0.05751780 0.2184029
<- c()
beta for(c in 1:C_true) for(n in 1:N[c])
<- cbind(beta, rmvnorm(mu = b_true[,c,drop=F], Sigma = matrix(Omega_true[,c,drop=F], ncol = P_r)))
beta <- rep(1:3, times = N) z_true
```

We specify the following prior parameters (for their definition see the vignette on model fitting):

```
<- 0.1
delta <- numeric(P_r)
xi <- diag(P_r)
D <- P_r + 2
nu <- diag(P_r) Theta
```

Initially, we start with `C = 1`

latent classes. The class
mean `b`

is set to zero, the covariance matrix
`Omega`

to the identity matrix:

```
<- rep(1, ncol(beta))
z <- 1
C <- matrix(0, nrow = P_r, ncol = C)
b <- matrix(rep(diag(P_r), C), nrow = P_r*P_r, ncol = C) Omega
```

The following call to `update_classes_dp()`

updates the
latent classes in `100`

iterations. Note that we specify the
arguments `Cmax`

and `s_desc`

. The former denotes
the maximum number of latent classes. This specification is not a
requirement for the Dirichlet process per se, but rather for its
implementation. Knowing the maximum possible class number, we can
allocate the required memory space, which leads to a speed improvement.
We later can verify that we won’t exceed the number of
`Cmax = 10`

latent classes at any point of the Dirichlet
process. Setting `s_desc = TRUE`

ensures that the classes are
ordered by their weights in a descending order to ensure
identifiability.

```
for(r in 1:100){
<- RprobitB:::update_classes_dp(
dp Cmax = 10, beta, z, b, Omega, delta, xi, D, nu, Theta, s_desc = TRUE
)<- dp$z
z <- dp$b
b <- dp$Omega
Omega }
```

The Dirichlet process was able to infer the true number
`C_true = 3`

of latent classes:

```
par(mfrow = c(1,2))
plot(t(beta), xlab = bquote(beta[1]), ylab = bquote(beta[2]), pch = 19)
:::plot_class_allocation(beta, z, b, Omega, r = 100, perc = 0.95) RprobitB
```

Aldous, D. J. 1985. “Exchangeability and Related Topics,”
Lecture notes in math., 1117: 1–198.

Bauer, D., S. Büscher, and M. Batram. 2019. “Non-Parameteric
Estiation of Mixed Discrete Choice Models.” *Second
International Choice Modelling Conference in Kobe*.

Blackwell, D., and J. MacQueen. 1973. “Ferguson Distributions via
Polya Urn Schemes.” *The Annals of Statistics* 1: 353–55.

Burda, M., M. Harding, and J. Hausman. 2008. “A Bayesian Mixed
Logit–Probit Model for Multinomial Choice.” *Journal of
Econometrics* 147 (2). https://doi.org/10.1016/j.jeconom.2008.09.029.

Croissant, Y. 2020. “Estimation of Random Utility Models in r: The
Mlogit Package.” *Journal of Statistical Software* 95
(11). https://doi.org/10.18637/jss.v095.i11.

Li, Y., E. Schofield, and M. Gönen. 2019. “A Tutorial on Dirichlet
Process Mixture Modeling.” *Journal of Mathematical
Psychology* 91. https://doi.org/10.1016/j.jmp.2019.04.004.

Neal, R. M. 2000. “Markov Chain Sampling Methods for Dirichlet
Process Mixture Models.” *Journal of Computational and
Graphical Statistics* 9 (2). https://doi.org/10.2307/1390653.

Rasmussen, C. E. 2000. *The Infinite Gaussian Mixture Model*.
Vol. 12. MIT Press.

Sethuraman, J. 1994. “A Constructive Definition of Dirichlet
Priors.” *Statistica Sinica* 4 (2): 639–50. http://www.jstor.org/stable/24305538.

This vignette is built using R 4.2.1 with the {RprobitB} 1.1.1 package.↩︎

Setting

`cor = FALSE`

instead returns the estimated covariance matrix.↩︎If either

`weight_update = TRUE`

or`dp_update = TRUE`

(i.e. if classes are updated),`C`

equals the initial number of latent classes.↩︎A similar approach in the context of discrete choice can be found in Burda, Harding, and Hausman (2008), where the Dirichlet process is applied to estimate a mixed logit-probit model.↩︎