glm.Rmd
The bamlss package is perfectly suitable for estimating
(Bayesian) generalized linear models (GLM) and provides infrastructures
for the estimation using very large data sets, too. Within the main
model fitting function bamlss()
, the possible
family
specifications for fitting GLMs are:
"gaussian"
or gaussian_bamlss()
,"beta"
or beta_bamlss()
,"binomial"
or binomial_bamlss()
,"gamma"
or gamma_bamlss()
,"poisson"
or poisson_bamlss()
.In addition, there is a wrapper function for the family objects
provided by the gamlss package (bamlss:Stasinopoulos+Rigby:2018?),
so in principle all gamlss families can be used by just passing
them to the family
argument in the bamlss()
function (see also the count data regression
example).
First, we illustrate how to fit standard GLMs and how to do inference using the bamlss framework. Aftwards, we show how to estimate GLMs using very large data set.
This example is taken from the AER package (Kleiber and Zeileis 2018, 2008) and is about labor force participation (yes/no) of women in Switzerland 1981. The data can be loaded with
## participation income age education youngkids oldkids foreign
## 1 no 10.78750 3.0 8 1 1 no
## 2 yes 10.52425 4.5 8 0 1 no
## 3 no 10.96858 4.6 9 0 0 no
## 4 no 11.10500 3.1 11 2 0 no
## 5 no 11.10847 4.4 12 0 2 no
## 6 yes 11.02825 4.2 12 0 1 no
The data frame contains of 872 observations on 7 variables, where
some of them might have a nonlinear influence on the response labor
participation
. Now, a standard Bayesian binomial logit
model can be fitted with
library("bamlss")
## First, set the seed for reproducibly.
set.seed(123)
## Model formula.
f <- participation ~ income + age + education + youngkids + oldkids + foreign + I(age^2)
## Estimate model.
b <- bamlss(f, family = "binomial", data = SwissLabor,
n.iter = 12000, burnin = 2000, thin = 10)
Note, to capture nonlinearities, a quadratic term for variable
age
is added to the model. The model summary gives
summary(b)
##
## Call:
## bamlss(formula = f, family = "binomial", data = SwissLabor, n.iter = 12000,
## burnin = 2000, thin = 10)
## ---
## Family: binomial
## Link function: pi = logit
## *---
## Formula pi:
## ---
## participation ~ income + age + education + youngkids + oldkids +
## foreign + I(age^2)
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) 6.29460 1.61320 6.35800 10.86894 6.196
## income -1.11826 -1.53526 -1.11857 -0.66780 -1.104
## age 3.45820 2.14132 3.45310 4.72934 3.437
## education 0.03269 -0.02513 0.03389 0.08447 0.033
## youngkids -1.17482 -1.50360 -1.17595 -0.83496 -1.186
## oldkids -0.24017 -0.39322 -0.24097 -0.07412 -0.241
## foreignyes 1.16362 0.77613 1.16502 1.53449 1.168
## I(age^2) -0.48985 -0.64783 -0.48968 -0.32830 -0.488
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.8684 0.2947 0.9857 1
## ---
## Sampler summary:
## -
## DIC = 1032.965 logLik = -512.6361 pd = 7.693
## runtime = 12.1
## ---
## Optimizer summary:
## -
## AICc = 1033.737 edf = 8 logLik = -508.7851
## logPost = -571.3986 nobs = 872 runtime = 0.171
which suggests “significant” effects for all covariates, since there are no zeros within the 95% credible intervals. Before we proceed, we usually do some convergence checks of the MCMC chains using traceplots.
plot(b, which = "samples")
The plots indicate convergence of the MCMC chains, i.e., there is no
visible trend in the MCMC chains and very low autocorrelation suggest
i.i.d. sampling from the posterior distribution. Note that the function
call would show all traceplots, however, for convenience we only show
the plots for the intercept and variable income
.
The estimated regression coefficients can also be extracted using the
coef()
method
## Extract posterior mean.
coef(b, FUN = mean)
## pi.p.(Intercept) pi.p.income pi.p.age pi.p.education
## 6.29459725 -1.11826351 3.45820448 0.03268852
## pi.p.youngkids pi.p.oldkids pi.p.foreignyes pi.p.I(age^2)
## -1.17481773 -0.24017030 1.16362217 -0.48984985
## pi.p.alpha
## 0.86841423
## Or use any other function on the samples.
coef(b, FUN = function(x) { quantile(x, prob = c(0.025, 0.975)) })
## 2.5% 97.5%
## pi.p.(Intercept) 1.61320356 10.86894390
## pi.p.income -1.53525827 -0.66779610
## pi.p.age 2.14132459 4.72934336
## pi.p.education -0.02513391 0.08446961
## pi.p.youngkids -1.50359765 -0.83496378
## pi.p.oldkids -0.39321512 -0.07412275
## pi.p.foreignyes 0.77612963 1.53449289
## pi.p.I(age^2) -0.64783455 -0.32830179
## pi.p.alpha 0.29466484 1.00000000
(there is also a confint()
method). The naming
convention of the regression coefficients might first seem a bit
atypical, it is based on the idea that a bamlss family/response
distribution can have more than just one distributional parameter, as
well as linear and/or nonlinear model terms. To explain, pi
is the name of the distributional parameter in the
binomial_bamlss()
family and p
stands for
parametric terms, i.e., there would also be a s
for smooth
terms if they are part of the model.
Model predictions on the probability scale can be obtained by the
predict method (see also function predict.bamlss()
).
## Create some newdata for prediction, note that
## factors need to be fully specified (this will be changed soon).
nd <- data.frame(income = 11, age = 3.3,
education = 12, youngkids = 1, oldkids = 1,
foreign = factor(1, levels = 1:2, labels = c("no", "yes")))
## Predicted probabilities.
predict(b, newdata = nd, type = "parameter")
## [1] 0.279716
nd$foreign <- "yes"
predict(b, newdata = nd, type = "parameter")
## [1] 0.552208
To visualize the effect of age on the probability we can do the following:
## Predict effect of age including 95% credible intervals and plot.
nd <- data.frame(income = 11, age = seq(2, 6.2, length = 100),
education = 12, youngkids = 1, oldkids = 1,
foreign = factor(1, levels = 1:2, labels = c("no", "yes")))
nd$pSwiss <- predict(b, newdata = nd, type = "parameter", FUN = c95)
nd$foreign <- "yes"
nd$pForeign <- predict(b, newdata = nd, type = "parameter", FUN = c95)
plot2d(p.no ~ age, data = nd, ylab = "participation",
ylim = range(c(nd$p.no, nd$pForeign)), lty = c(2, 1, 2))
plot2d(pForeign ~ age, data = nd, col.lines = "blue", add = TRUE,
lty = c(2, 1, 2))
legend("topright", c("foreign yes", "foreign no"), lwd = 1,
col = c("blue", "black"))
The plot nicely depicts the nonlinear effect of variable
age
for the two levels of foreign
.
This example is taken from Zeileis, Kleiber, and Jackman (2008). The application is about modeling the demand for medical care by elderly and was first analyzed by Deb and Trivedi (1997). The data can be downloaded on loaded into R with
download.file(
"https://www.jstatsoft.org/index.php/jss/article/downloadSuppFile/v027i08/DebTrivedi.rda.zip",
"DebTrivedi.rda.zip"
)
unzip("DebTrivedi.rda.zip")
load("DebTrivedi.rda")
head(DebTrivedi)
## ofp ofnp opp opnp emer hosp health numchron adldiff region age black gender
## 1 5 0 0 0 0 1 average 2 no other 6.9 yes male
## 2 1 0 2 0 2 0 average 2 no other 7.4 no female
## 3 13 0 0 0 3 3 poor 4 yes other 6.6 yes female
## 4 16 0 5 0 1 1 poor 2 yes other 7.6 no male
## 5 3 0 0 0 0 0 average 2 yes other 7.9 no female
## 6 17 0 0 0 0 0 poor 5 yes other 6.6 no female
## married school faminc employed privins medicaid
## 1 yes 6 2.8810 yes yes no
## 2 yes 10 2.7478 no yes no
## 3 no 10 0.6532 no no yes
## 4 yes 3 0.6588 no yes no
## 5 yes 6 0.6588 no yes no
## 6 no 7 0.3301 no no yes
The response variable in this data is the number of physician office
visits, variable ofp
. A histogram of the count data
response can be plotted with
which shows very large counts and also a large number of zero counts. Our first model is a Poisson regression, which can be estimated with
## Set the seed for reproducibly.
set.seed(123)
## Model formula.
f <- ofp ~ hosp + health + numchron + gender + school + privins
## Estimate model.
b1 <- bamlss(f, family = "poisson", data = DebTrivedi,
n.iter = 12000, burnin = 2000, thin = 10)
The model summary is shown by
summary(b1)
##
## Call:
## bamlss(formula = f, family = "poisson", data = DebTrivedi, n.iter = 12000,
## burnin = 2000, thin = 10)
## ---
## Family: poisson
## Link function: lambda = log
## *---
## Formula lambda:
## ---
## ofp ~ hosp + health + numchron + gender + school + privins
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) 1.03049 0.98540 1.03074 1.07790 1.029
## hosp 0.16452 0.15322 0.16440 0.17678 0.165
## healthpoor 0.24810 0.21558 0.24791 0.27968 0.248
## healthexcellent -0.36253 -0.42490 -0.36207 -0.30116 -0.362
## numchron 0.14669 0.13829 0.14641 0.15585 0.147
## gendermale -0.11279 -0.13677 -0.11268 -0.08745 -0.112
## school 0.02606 0.02223 0.02611 0.02960 0.026
## privinsyes 0.20102 0.16795 0.20116 0.23446 0.202
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.9644 0.8023 0.9986 1
## ---
## Sampler summary:
## -
## DIC = 35958.86 logLik = -17975.52 pd = 7.813
## runtime = 34.156
## ---
## Optimizer summary:
## -
## AICc = 35959.26 edf = 8 logLik = -17971.61
## logPost = -18034.23 nobs = 4406 runtime = 0.241
which shows “significant” effects for all covariates, since there are no zeros within the 95% credible intervals. We can use randomized quantile residuals (Dunn and Smyth 1996) to evaluate how good the models fits to the data.
## Warning in residuals.bamlss(x, FUN = FUN, ...): non finite quantiles from
## probabilities, set to NA!
Clearly, the plots shows that the model performance is not very good at the tails of the data, this might be caused by the large amount of zero counts in the data, which is not accounted for using the Poisson distribution.
Therefore, another possible candidate for the response distribution
is the zero-inflated negative binomial distribution, which is
implemented in the gamlss package (bamlss:Stasinopoulos+Rigby:2018?)
in the function ZINBI
. The family has 3 parameters
mu
, sigma
and nu
. In this example
we model distributional parameters mu
and
sigma
in terms of covariates by setting up the model
formula with
f <- list(
ofp ~ hosp + health + numchron + gender + school + privins,
sigma ~ hosp + health + numchron + gender + school + privins,
nu ~ 1
)
Note, for parameter nu
an intercept only model would
also be the default if the specification is omitted in the model
formula. The model is estimated with (note, this can take some time)
## Get the gamlss families.
library("gamlss")
## Set the seed for reproducibly.
set.seed(123)
## Estimate model.
b2 <- bamlss(f, family = ZINBI, data = DebTrivedi,
n.iter = 12000, burnin = 2000, thin = 10)
The model summary gives
summary(b2)
##
## Call:
## bamlss(formula = f, family = ZINBI, data = DebTrivedi, n.iter = 12000,
## burnin = 2000, thin = 10)
## ---
## Family: ZINBI
## Link function: mu = log, sigma = log, nu = logit
## *---
## Formula mu:
## ---
## ofp ~ hosp + health + numchron + gender + school + privins
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) 1.01470 0.88866 1.01485 1.13262 1.016
## hosp 0.20735 0.16496 0.20773 0.24950 0.210
## healthpoor 0.21927 0.12495 0.22063 0.31016 0.221
## healthexcellent -0.35063 -0.47696 -0.34898 -0.23186 -0.341
## numchron 0.15875 0.13652 0.15845 0.18334 0.159
## gendermale -0.11469 -0.17365 -0.11470 -0.05689 -0.115
## school 0.02641 0.01720 0.02630 0.03495 0.027
## privinsyes 0.18738 0.09794 0.18607 0.27407 0.184
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.67619 0.08023 0.71540 1
## ---
## Formula sigma:
## ---
## sigma ~ hosp + health + numchron + gender + school + privins
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) 0.510961 0.316771 0.508389 0.706571 0.520
## hosp -0.028914 -0.095758 -0.028909 0.038187 -0.023
## healthpoor 0.227347 0.063571 0.231602 0.396216 0.227
## healthexcellent -0.002851 -0.228342 -0.007384 0.234369 0.003
## numchron -0.236836 -0.278745 -0.236980 -0.194826 -0.236
## gendermale 0.222175 0.105449 0.223875 0.332764 0.225
## school -0.012696 -0.028136 -0.012729 0.003863 -0.013
## privinsyes -0.449505 -0.587264 -0.452838 -0.308850 -0.452
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.61640 0.05323 0.63030 1
## ---
## Formula nu:
## ---
## nu ~ 1
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) -4.210 -5.071 -4.177 -3.607 -4.268
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.9091 0.4316 0.9868 1
## ---
## Sampler summary:
## -
## DIC = 24206.2 logLik = -12094.4 pd = 17.3959
## runtime = 987.661
## ---
## Optimizer summary:
## -
## AICc = 24205.35 edf = 17 logLik = -12085.61
## logPost = -12218.66 nobs = 4406 runtime = 2.508
and the corresponding randomized quantile residuals plots are created with
The plots indicate a better model fit compared to the Poisson model, however, for very large counts the model performance could probably be improved further. The corresponding DIC values are
DIC(b1, b2)
## DIC pd
## b1 35958.86 7.813035
## b2 24206.20 17.395871
which also state that the zero-inflated negative binomial model is better than the Poisson model.
The R package bamlss provides infrastructures to estimate
models using large data sets. The default algorithms for posterior mode
estimation (see function opt_bfit()
) and MCMC simulation
(function sam_GMCMC()
) can be utilized in these situations.
More precisely, the algorithms make advantage of the fact that usually
the number of unique observations is much smaller than the number of
total observations in the data (for a detailed description of the
algorithms see Lang et al. (2014)). For
illustration, we load the precipitation data of the HOMSTART-project
which was first analyzed in Umlauf et al.
(2012) (note, an internet connection is required).
homstart_data(load = TRUE)
head(homstart)
## raw cens bin cat trend month year day lon lat id cos1
## 1 NA NA <NA> <NA> -22.00000 Jan 1948 1 14.75 47.6 1 0.9998518
## 2 37.0 37.0 yes high -21.99726 Jan 1948 2 14.75 47.6 1 0.9994074
## 3 3.0 3.0 yes medium -21.99452 Jan 1948 3 14.75 47.6 1 0.9986668
## 4 0.0 0.0 no none -21.99178 Jan 1948 4 14.75 47.6 1 0.9976303
## 5 5.2 5.2 yes high -21.98904 Jan 1948 5 14.75 47.6 1 0.9962982
## 6 0.0 0.0 no none -21.98630 Jan 1948 6 14.75 47.6 1 0.9946708
## cos2 sin1 sin2 weekend elevation
## 1 0.9994074 0.01721336 0.03442161 no 779
## 2 0.9976303 0.03442161 0.06880243 no 779
## 3 0.9946708 0.05161967 0.10310170 yes 779
## 4 0.9905325 0.06880243 0.13727877 yes 779
## 5 0.9852201 0.08596480 0.17129314 no 779
## 6 0.9787401 0.10310170 0.20510450 no 779
dim(homstart)
## [1] 1063610 17
The data consists of 1063610 observations. To illustrate how large
models can be fitted using bamlss, we estimate a logit model
using the binary response variable bin
, which indicates if
it rained or not.
## Set the seed for reproducibly.
set.seed(123)
## Model formula including a seasonal effect of the day
## of the year using harmonic regression.
f <- bin ~ weekend + elevation + cos1 + cos2 + sin1 + sin2
## Estimate model using only unique observations by setting
## argument binning = TRUE in the bamlss() call. Moreover,
## only use minimum storage for the returned object by
## setting light = TRUE.
b <- bamlss(f, data = homstart, family = "binomial",
binning = TRUE, light = TRUE)
By setting binning = TRUE
, the algorithms only use 33580
unique observations for model fitting. This could in principle even be
further utilized if each covariate represents one model term, e.g.:
f <- bin ~ weekend +
s(elevation,bs="re",fx=TRUE) +
s(cos1,bs="re",fx=TRUE) + s(cos2,bs="re",fx=TRUE) +
s(sin1,bs="re",fx=TRUE) + s(sin2,bs="re",fx=TRUE)
Which is a bit of a hack using the random effects smooth constructor
and setting fx = TRUE
, forcing the smoother to use fixed
effects (linear effects) estimation. However, by doing so, the resulting
number of unique observations for the harmonic part is only 365 and for
elevation only 46, which saves quite a lot of memory storage.
In this example, we are interested if it is more likely to rain on weekends than during the week (see also Umlauf et al. (2012) and Umlauf, Klein, and Zeileis (2018) for a more detailed analysis). The model summary suggests
summary(b)
##
## Call:
## bamlss(formula = f, family = "binomial", data = homstart, light = TRUE,
## binning = TRUE)
## ---
## Family: binomial
## Link function: pi = logit
## *---
## Formula pi:
## ---
## bin ~ weekend + elevation + cos1 + cos2 + sin1 + sin2
## -
## Parametric coefficients:
## Mean 2.5% 50% 97.5% parameters
## (Intercept) -3.118e-01 -3.209e-01 -3.118e-01 -3.028e-01 -0.312
## weekendyes 1.241e-02 4.298e-03 1.244e-02 2.083e-02 0.012
## elevation 6.440e-05 5.241e-05 6.433e-05 7.664e-05 0.000
## cos1 -2.652e-01 -2.707e-01 -2.652e-01 -2.596e-01 -0.265
## cos2 1.462e-01 1.405e-01 1.461e-01 1.520e-01 0.146
## sin1 7.021e-02 6.494e-02 7.025e-02 7.590e-02 0.070
## sin2 -2.897e-02 -3.462e-02 -2.900e-02 -2.359e-02 -0.029
## -
## Acceptance probability:
## Mean 2.5% 50% 97.5%
## alpha 0.9988 0.9918 1.0000 1
## ---
## Sampler summary:
## -
## runtime = 532.388
## ---
## Optimizer summary:
## -
## AICc = 1330320 edf = 7 logLik = -665152.9
## logPost = -665207.7 nobs = 980633 runtime = 2.349
that the weekend effect is very little and since we did not account
for any spatial correlations in this example, the effect is most
probably not existing (at least in Austria). Note, because we set
light = TRUE
in the bamlss()
call, there will
be no real sampler summary except of the presented runtime of the
sampler. The reason is, that computing the DIC from the posterior
samples can take quite long and can lead to memory storage problems,
too.