# Summer Break

The blog will be off for a few days… probably weeks. I am currently packing with the family, since we’re moving…

I will be back in September….

# Multivariate and dynamic risk measures

After a few years, I decided to put online some lectures notes I had from a graduate course I gave over one (long) day in 2014, in Leuven, entitled “an introduction to multivariate and dynamic risk measures”. The notes are now available on hal. I just hope that it might be usefull to someone…

# On the robustness of LASSO

Probably the last post on lasso, before the summer break… More specifically, I was wondering about the interpretation of graphs $\lambda\mapsto\widehat{\beta}_\lambda$. We use them for variable selection, but my major concern was about confidence intervals : how can we trust those lines ?

As usual, a natural way is to use simulations on generated datasets. Consider for instance

Sigma = matrix(c(1,.8,.2,.8,1,.4,.2,.4,1),3,3) n = 1000 library(mnormt) X = rmnorm(n,rep(0,3),Sigma) set.seed(123) df = data.frame(X1=X[,1],X2=X[,2],X3=X[,3],X4=rnorm(n), X5=runif(n), X6=exp(X[,3]), X7=sample(c("A","B"),size=n,replace=TRUE,prob=c(.5,.5)), X8=sample(c("C","D"),size=n,replace=TRUE,prob=c(.5,.5))) df$Y = 1+df$X1-df$X4+5*(df$X7=="A")+rnorm(n)

One can use other simulations of datasets, and store the output

vlambda = exp(seq(-8,1,length=201)) lasso = glmnet(x=X,y=df[,"Y"],family="gaussian",alpha=1, lambda=vlambda,standardize=TRUE) VLASSO[[s]] = as.matrix(lasso$beta) To visualize confidence bands, one can compute quantiles Q05=Q95=Qm=matrix(NA,9,201) for(i in 1:nrow(Q05)){ for(j in 1:ncol(Q05)){ v = unlist(lapply(VLASSO,function(x) x[i,j])) Q05[i,j] = quantile(v,.05) Q95[i,j] = quantile(v,.95) Qm[i,j] = mean(v) }} and get get the graph plot(lasso,col=colrs,"lambda"ylim=c(min(Q05),max(Q95))) colrs=c(brewer.pal(8,"Set1")) polygon(c(log(lasso$lambda),rev(log(lasso$lambda))), c(Q05[2,],rev(Q95[2,])),col=colrs[1],border=NA) polygon(c(log(lasso$lambda),rev(log(lasso$lambda))), c(Q05[5,],rev(Q95[5,])),col=colrs[2],border=NA) polygon(c(log(lasso$lambda),rev(log(lasso$lambda))), c(Q05[8,],rev(Q95[8,])),col=colrs[3],border=NA) An alternative (more realistic on real data) is to use bootstrapped version of the dataset id = sample(1:nrow(X),size=nrow(X),replace=TRUE) lasso = glmnet(x=X[id,],y=df[id,"Y"],family="gaussian",alpha=1, lambda=vlambda,standardize=TRUE) So far, it looks it’s working very well. Now, what if we have a smaller dataset n = 100 On simulated new samples, we get while the bootstrap version is There is more uncertainty, clearly, but the conclusion is not ambiguous here. Now, what about real data. Consider the following chicago = read.table("http://freakonometrics.free.fr/chicago.txt",header=TRUE,sep=";") tail(chicago) Fire X_1 X_2 X_3 42 4.8 0.152 19 13.323 43 10.4 0.408 25 12.960 44 15.6 0.578 28 11.260 45 7.0 0.114 3 10.080 46 7.1 0.492 23 11.428 47 4.9 0.466 27 13.731 with one variable of interest (the number of fires, per unhabitants) and 3 features. We can here use bootstrap to generate samples, and then fit a lasso regression. On the original dataset, the regression is X = model.matrix(lm(Fire~.,data=chicago)) id = sample(1:nrow(X),size=nrow(X),replace=TRUE) vlambda = exp(seq(-4,2,length=201)) lasso = glmnet(x=X[id,],y=chicago[id,"Fire"],family="gaussian",alpha=1, lambda=vlambda,standardize=TRUE) And if we just plot lines $\lambda\mapsto\widehat{\beta}_\lambda$ we get Now, consider bootstrap samples. for(s in 1:100){ id=sample(1:nrow(X),size=nrow(X),replace=TRUE) library(glmnet) vlambda=exp(seq(-4,2,length=201)) lasso=glmnet(x=X[id,],y=chicago[id,"Fire"],family="gaussian",alpha=1, lambda=vlambda,standardize=TRUE) plot(lasso,col=colrs,"lambda",lwd=.2,add=TRUE)} We get here The interpretation here is much more difficult What about the order ? N=matrix(NA,100000,4) for(s in 1:100000){ id=sample(1:nrow(X),size=nrow(X),replace=TRUE) library(glmnet) vlambda=exp(seq(-4,2,length=201)) lasso=glmnet(x=X[id,],y=chicago[id,"Fire"], family="gaussian",alpha=1, lambda=vlambda,standardize=TRUE) N[s,]=names(sort(apply(as.matrix(lasso$beta), 1,function(x) sum(x!=0))))}

The ordering that was obtained on the original dataset was the same in 56% of the scenarios,

mean(apply(N,1,function(x) paste(x,collapse="")=="(Intercept)X_1X_2X_3")) [1] 0.5693

We can look at all the cases,

L=as.character(c(123,132,213,231,312,321)) Li=paste("(Intercept)X_",substr(L,1,1),"X_", substr(L,2,2),"X_",substr(L,3,3),sep="") g=function(y) mean(apply(N,1,function(x) paste(x,collapse="")==y)) vL=unlist(lapply(Li,g)) names(vL)=L barplot(vL,las=2,horiz=TRUE)

# Standardization in LASSO

The lasso regression is based on the idea of solving$$\widehat{\mathbf{\beta}}_{\lambda}=\text{argmin}\lbrace -\log\mathcal{L}(\mathbf{\beta}|\mathbf{x},\mathbf{y})+\lambda\|\mathbf{\beta}\|_{\ell_1}\rbrace$$where$$\Vert\mathbf{a} \Vert_{\ell_1}=\sum_{i=1}^d |a_i|$$for any $\mathbf{a}\in\mathbb{R}^d$. In a recent post, we’ve seen computational aspects of the optimization problem. But I went quickly throught the story of the $\ell_1$-norm. Because it means, somehow, that the value of $\beta_1$ and $\beta_2$ should be comparable. Somehow, with two significant variables, with very different scales, we should expect orders (or relative magnitudes) of $\widehat{\beta}_1$ and $\widehat{\beta}_2$ to be very very different. So people say that it is therefore necessary to center and reduce (or standardize) the variables.

Consider the following (simulated) dataset

Sigma = matrix(c(1,.8,.2,.8,1,.4,.2,.4,1),3,3) n = 1000 library(mnormt) X = rmnorm(n,rep(0,3),Sigma) set.seed(123) df = data.frame(X1=X[,1],X2=X[,2],X3=X[,3],X4=rnorm(n), X5=runif(n),X6=exp(X[,3]), X7=sample(c("A","B"),size=n,replace=TRUE,prob=c(.5,.5)), X8=sample(c("C","D"),size=n,replace=TRUE,prob=c(.5,.5))) df$Y = 1+df$X1-df$X4+5*(df$X7=="A")+rnorm(n) X = model.matrix(lm(Y~.,data=df))

Use the following colors for the graphs and the value of $\lambda$

library("RColorBrewer") colrs = c(brewer.pal(8,"Set1"))[c(1,4,5,2,6,3,7,8)] vlambda=exp(seq(-8,1,length=201))

The first regression we can run is a non-standardized one

library(glmnet) lasso = glmnet(x=X,y=df[,"Y"],family="gaussian",alpha=1,lambda=vlambda,standardize=FALSE)

We can visualize the graphs of $\lambda\mapsto\widehat{\beta}_\lambda$

idx = which(apply(lasso$beta,1,function(x) sum(x==0))&lt;200) plot(lasso,col=colrs,'lambda',xlim=c(-5.5,2.3),lwd=2) legend(1.2,.9,legend=paste('X',0:8,sep='')[idx],col=colrs,lty=1,lwd=2) At least, observe that the most significant variables are the one that were used to generate the data. Now, consider the case that we standardize the data lasso = glmnet(x=X,y=df[,"Y"],family="gaussian",alpha=1,lambda=vlambda,standardize=TRUE) The graphs of $\lambda\mapsto\widehat{\beta}_\lambda$ The graph is (strangely) very similar to the previous one. Except perhaps for the green curve. Maybe that categorical are not simular to continuous variables… Because somehow, standardisation of categorical variables might be not natural… Why not consider some home-made function ? Let us transform (linearly) all variable in the $X$ matrix (except the first one, which is the intercept) Xc = X for(j in 2:ncol(X)) Xc[,j]=(Xc[,j]-mean(Xc[,j]))/sd(Xc[,j]) Now, we can run our lasso regression on that one (with the intercept since all the variables are centered, but $y$) lasso = glmnet(x=Xc,y=df$Y,family="gaussian",alpha=1,intercept=TRUE,lambda=vlambda)

The plot is now

plot(lasso,col=colrs,"lambda",xlim=c(-6.7,1.3),lwd=2) idx = which(apply(lasso$beta,1,function(x) sum(x==0))&lt;length(vlambda)) legend(.15,.45,legend=paste('X',0:8,sep='')[idx],col=colrs,lty=1,bty=&quot;n&quot;,lwd=2) Actually, why not also center the $y$ variable, and remove also the intercept Yc = (df[,"Y"]-mean(df[,"Y"]))/sd(df[,"Y"]) lasso = glmnet(x=Xc,y=Yc,family="gaussian",alpha=1,intercept=FALSE,lambda=vlambda) Hopefully, those graphs are very consistent (and if we use those for variable selection, they suggest to use variables that were actually used to generate the dataset). And having qualitative and quantitative variable is not a big deal. But still, I do not feel confortable with the differences… # Short Break in Barcelona Now that the course is over, and that I have been to the Biometrics conference, I will enjoy a short break with the kids in Barcelona… I will be off. Completely. # Biometrics Conference, Barcelona This week, I will be at the XXIX International Biometric conference, in Barcelona, to give a talk on massive collaborative data to study mortality (in an invited session, on Tuesday afternoon). Slides are available online. # Convex Regression Model This morning during the lecture on nonlinear regression, I mentioned (very) briefly the case of convex regression. Since I forgot to mention the codes in R, I will publish them here. Assume that $y_i=m(\mathbf{x}_i)+\varepsilon_i$ where $m:\mathbb{R}^d\rightarrow \mathbb{R}$ is some convex function. Then $m$ is convex if and only if $\forall\mathbf{x}_1,\mathbf{x}_2\in\mathbb{R}^d$, $\forall t\in[0,1]$, $$m(t\mathbf{x}_1+[1-t]\mathbf{x}_2) \leq tm(\mathbf{x}_1)+[1-t]m(\mathbf{x}_2)$$Hidreth (1954) proved that if$$m^\star=\underset{m \text{ convex}}{\text{argmin}}\left\lbrace\sum_{i=1}^n \big(y_i-m(\mathbf{x_i})\big)^2\right\rbrace$$then $\mathbf{\theta}^\star=(m^\star(\mathbf{x_1}),\cdots,m^\star(\mathbf{x_n}))$ is unique. Let $\mathbf{y}=\mathbf{\theta}+\mathbf{\varepsilon}$, then $$\mathbf{\theta}^\star=\underset{\mathbf{\theta}\in \mathcal{K}}{\text{argmin}}\left\lbrace\sum_{i=1}^n \big(y_i-\theta_i)\big)^2\right\rbrace$$where$$\mathcal{K}=\{\mathbf{\theta}\in\mathbb{R}^n:\exists m\text{ convex },m(\mathbf{x}_i)=\theta_i\}$$. I.e. $\mathbf{\theta}^\star$ is the projection of $\mathbf{y}$ onto the (closed) convex cone $\mathcal{K}$. The projection theorem gives existence and unicity. For convenience, in the application, we will consider the real-valued case, $m:\mathbb{R}\rightarrow \mathbb{R}$, i.e. $y_i=m(x_i)+\varepsilon_i$. Assume that observations are ordered $x_1\leq x_2\leq\cdots \leq x_n$. Here $$\mathcal{K}=\left\lbrace\mathbf{\theta}\in\mathbb{R}^n:\frac{\theta_2-\theta_1}{x_2-x_1}\leq \frac{\theta_3-\theta_2}{x_3-x_2}\leq \cdots \leq \frac{\theta_n-\theta_{n-1}}{x_n-x_{n-1}}\right\rbrace$$ Hence, quadratic program with $n-2$ linear constraints. $m^\star$ is a piecewise linear function (interpolation of consecutive pairs $(x_i,\theta_i^\star)$). If $m$ is differentiable, $m$ is convex if $$m(\mathbf{x})+ \nabla m(\mathbf{x})^{\text{T}}\cdot[\mathbf{y}-\mathbf{x}] \leq m(\mathbf{y})$$ More generally, if $m$ is convex, then there exists $\xi_{\mathbf{x}}\in\mathbb{R}^n$ such that $$m(\mathbf{x})+ \xi_{\mathbf{x}}^{\text{ T}}\cdot[\mathbf{y}-\mathbf{x}] \leq m(\mathbf{y})$$ $\xi_{\mathbf{x}}$ is a subgradient of $m$ at ${\mathbf{x}}$. And then $$\partial m(\mathbf{x})=\big\lbrace m(\mathbf{x})+ \xi^{\text{ T}}\cdot[\mathbf{y}-\mathbf{x}] \leq m(\mathbf{y}),\forall \mathbf{y}\in\mathbb{R}^n\big\rbrace$$ Hence, $\mathbf{\theta}^\star$ is solution of $$\text{argmin}\big\lbrace\|\mathbf{y}-\mathbf{\theta}\|^2\big\rbrace$$$$\text{subject to }\theta_i+\xi_i^{\text{ T}}[\mathbf{x}_j-\mathbf{x}_i]\leq\mathbf{\theta}_j,~\forall i,j$$ and $\xi_1,\cdots,\xi_n\in\mathbb{R}^n$. Now, to do it for real, use cobs package for constrained (b)splines regression, library(cobs) To get a convex regression, use plot(cars) x = cars$speed y = cars\$dist rc = conreg(x,y,convex=TRUE) lines(rc, col = 2)

Here we can get the values of the knots

rc   Call: conreg(x = x, y = y, convex = TRUE) Convex regression: From 19 separated x-values, using 5 inner knots, 7, 8, 9, 20, 23. RSS = 1356; R^2 = 0.8766; needed (5,0) iterations

and actually, if we use them in a linear-spline regression, we get the same output here

reg = lm(dist~bs(speed,degree=1,knots=c(4,7,8,9,,20,23,25)),data=cars) u = seq(4,25,by=.1) v = predict(reg,newdata=data.frame(speed=u)) lines(u,v,col="green")

Let us add vertical lines for the knots

abline(v=c(4,7,8,9,20,23,25),col="grey",lty=2)

# 7eme rencontres R

Cette fin de semaine, les 7emes rencontres R sont organisées a Rennes. Ewen fera une (courte) présentation vendredi matin de nos travaux en démographie… Les slides sont en ligne (et l’article aussi).

# Summer School, Big Data and Economics

This week I will be giving a lecture at the  2018 edition of the Summer School at the UB School of Economics, in Barcelona. It will be a four day crash course, starting on Tuesday (morning).

Lecture 1: Introduction : Why Big Data brings New Questions
Lecture 2: Simulation Based Techniques & Bootstrap
Lecture 3: Loss Functions : from OLS to Quantile Regression
Lecture 4: Nonlinearities and Discontinuities
Lecture 5: Cross-Validation and Out-of-Sample diagnosis
Lecture 6: Variable and model selection
Lecture 7: New Tools for Classification Problems
Lecture 8: New Tools for Time Series & Forecasting

Some slides are available on github, and probably more interesting, I will upload a R markdown with all the codes.