Tag Archives: convex

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\rbracethen \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\rbracewhere\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)

Classification from scratch, penalized Ridge logistic 4/8

Fourth post of our series on classification from scratch, following the previous post which was some sort of detour on kernels. But today, we’ll get back on the logistic model.

Formal approach of the problem

We’ve seen before that the classical estimation technique used to estimate the parameters of a parametric model was to use the maximum likelihood approach. More specifically, \widehat{\mathbf{\beta}}=\text{argmax}\lbrace \log\mathcal{L}(\mathbf{\beta}|\mathbf{x},\mathbf{y})\rbraceThe objective function here focuses (only) on the goodness of fit. But usually, in econometrics, we believe something like non sunt multiplicanda entia sine necessitate (“entities are not to be multiplied without necessity”), the parsimony principle, simpler theories are preferable to more complex ones. So we want to penalize for too complex models.

This is not a bad idea. It is mentioned here and there in econometrics textbooks, but usually, for model choice, not about the inference. Usually, we estimate parameters using maximum likelihood techniques, and them we use AIC or BIC to compare two models. Recall that Akaike (AIC) criteria is based on-2\log\mathcal{L}(\widehat{\mathbf{\beta}}|\mathbf{x},\mathbf{y})+2\text{dim}(\widehat{\mathbf{\beta}})We have on the left a measure for the goodness of fit, and on the right, a penalty increasing with the “complexity” of the model.

Very quickly, here, the complexity is the number of variates used. I will not enter into details about the concept of sparsity (and the true dimension of the problem), I will recommend to read the book by Martin Wainwright, Robert Tibshirani and Trevor Hastie on that issue. But assume that we do not make and variable selection, we consider the regression on all covariates. Define\Vert\mathbf{a} \Vert_{\ell_0}=\sum_{i=1}^d \mathbf{1}(a_i\neq 0), ~~\Vert\mathbf{a} \Vert_{\ell_1}=\sum_{i=1}^d |a_i|,~~\Vert\mathbf{a} \Vert_{\ell_2}=\left(\sum_{i=1}^d a_i^2\right)^{1/2}for any \mathbf{a}\in\mathbb{R}^d. One might say that the AIC could be written-2\log\mathcal{L}(\widehat{\mathbf{\beta}}|\mathbf{x},\mathbf{y})+2\|\widehat{\mathbf{\beta}}\|_{\ell_0}And actually, this will be our objective function. More specifically, we will consider
\widehat{\mathbf{\beta}}_{\lambda}=\text{argmin}\lbrace -\log\mathcal{L}(\mathbf{\beta}|\mathbf{x},\mathbf{y})+\lambda\|\mathbf{\beta}\|\rbracefor some norm \|\cdot\|. I will not get back here on the motivation and the (theoretical) properties of those estimates (that will actually be discussed in the Summer School in Barcelona, in July), but in this post, I want to discuss the numerical algorithm to solve such optimization problem, for \|\cdot\|_{\ell_2} (the Ridge regression) and for \|\cdot\|_{\ell_1} (the LASSO regression).

Normalization of the covariates

The problem of \|\mathbf{\beta}\| is that the norm should make sense, somehow. A small \mathbf{\beta}_j is with respect to the “dimension” of x_j‘s. So, the first step will be to consider linear transformations of all covariates x_j to get centered and scaled variables (with unit variance)

y = myocarde$PRONO
X = myocarde[,1:7]
for(j in 1:7) X[,j] = (X[,j]-mean(X[,j]))/sd(X[,j])
X = as.matrix(X)

Ridge Regression (from scratch)

Before running some codes, recall that we want to solve something like\widehat{\mathbf{\beta}}_{\lambda}=\text{argmin}\lbrace -\log\mathcal{L}(\mathbf{\beta}|\mathbf{x},\mathbf{y})+\lambda\|\mathbf{\beta}\|_{\ell_2}^2\rbrace In the case where we consider the log-likelihood of some Gaussian variable, we get the sum of the square of the residuals, and we can obtain an explicit solution. But not in the context of a logistic regression.

The heuristics about Ridge regression is the following graph. In the background, we can visualize the (two-dimensional) log-likelihood of the logistic regression, and the blue circle is the constraint we have, if we rewite the optimization problem as a contrained optimization problem : \min_{\mathbf{\beta}:\|\mathbf{\beta}\|^2_{\ell_2}\leq s} \lbrace \sum_{i=1}^n -\log\mathcal{L}(y_i,\beta_0+\mathbf{x}^T\mathbf{\beta}) \rbracecan be written equivalently (it is a strictly convex problem)\min_{\mathbf{\beta},\lambda} \lbrace -\sum_{i=1}^n \log\mathcal{L}(y_i,\beta_0+\mathbf{x}^T\mathbf{\beta}) +\lambda \|\mathbf{\beta}\|_{\ell_2}^2 \rbraceThus, the constrained maximum should lie in the blue disk

LogLik = function(bbeta){
  b0=bbeta[1]
  beta=bbeta[-1]
  sum(-y*log(1 + exp(-(b0+X%*%beta))) - 
  (1-y)*log(1 + exp(b0+X%*%beta)))}
u = seq(-4,4,length=251)
v = outer(u,u,function(x,y) LogLik(c(1,x,y)))
image(u,u,v,col=rev(heat.colors(25)))
contour(u,u,v,add=TRUE)
u = seq(-1,1,length=251)
lines(u,sqrt(1-u^2),type="l",lwd=2,col="blue")
lines(u,-sqrt(1-u^2),type="l",lwd=2,col="blue")

Let us consider the objective function, with the following code

PennegLogLik = function(bbeta,lambda=0){
  b0   = bbeta[1]
  beta = bbeta[-1]
 -sum(-y*log(1 + exp(-(b0+X%*%beta))) - (1-y)*
  log(1 + exp(b0+X%*%beta)))+lambda*sum(beta^2)
}

Why not try a standard optimisation routine ? In the very first post on that series, we did mention that using optimization routines were not clever, since they were strongly relying on the starting point. But here, it is not the case

lambda = 1
beta_init = lm(PRONO~.,data=myocarde)$coefficients
vpar = matrix(NA,1000,8)
for(i in 1:1000){
vpar[i,] = optim(par = beta_init*rnorm(8,1,2), 
function(x) PennegLogLik(x,lambda), method = "BFGS", control = list(abstol=1e-9))$par}
par(mfrow=c(1,2))
plot(density(vpar[,2]),ylab="",xlab=names(myocarde)[1])
plot(density(vpar[,3]),ylab="",xlab=names(myocarde)[2])


Clearly, even if we change the starting point, it looks like we converge towards the same value. That could be considered as the optimum.

The code to compute \widehat{\mathbf{\beta}}_{\lambda} would then be

opt_ridge = function(lambda){
beta_init = lm(PRONO~.,data=myocarde)$coefficients
logistic_opt = optim(par = beta_init*0, function(x) PennegLogLik(x,lambda), 
method = "BFGS", control=list(abstol=1e-9))
logistic_opt$par[-1]}

and we can visualize the evolution of \widehat{\mathbf{\beta}}_{\lambda} as a function of {\lambda}

v_lambda = c(exp(seq(-2,5,length=61)))
est_ridge = Vectorize(opt_ridge)(v_lambda)
library("RColorBrewer")
colrs = brewer.pal(7,"Set1")
plot(v_lambda,est_ridge[1,],col=colrs[1])
for(i in 2:7) lines(v_lambda,est_ridge[i,],col=colrs[i])

At least it seems to make sense: we can observe the shrinkage as \lambda increases (we’ll get back to that later on).

Ridge, using Netwon Raphson algorithm

We’ve seen that we can also use Newton Raphson to solve this problem. Without the penalty term, the algorithm was\mathbf{\beta}_{new} = \mathbf{\beta}_{old} - \left(\frac{\partial^2\log\mathcal{L}(\mathbf{\beta}_{old})}{\partial\mathbf{\beta}\partial\mathbf{\beta}^T}\right)^{-1}\cdot \frac{\partial\log\mathcal{L}(\mathbf{\beta}_{old})}{\partial\mathbf{\beta}}where
\frac{\partial\log\mathcal{L}(\mathbf{\beta}_{old})}{\partial\mathbf{\beta}}=\mathbf{X}^T(\mathbf{y}-\mathbf{p}_{old})and\frac{\partial^2\log\mathcal{L}(\mathbf{\beta}_{old})}{\partial\mathbf{\beta}\partial\mathbf{\beta}^T}=-\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{X}where \mathbf{\Delta}_{old} is the diagonal matrix with terms \mathbf{p}_{old}(1-\mathbf{p}_{old}) on the diagonal.

Thus\mathbf{\beta}_{new} = \mathbf{\beta}_{old} + (\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{X})^{-1}\mathbf{X}^T[\mathbf{y}-\mathbf{p}_{old}]that we can also write\mathbf{\beta}_{new} =(\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{X})^{-1}\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{z}where \mathbf{z}=\mathbf{X}\mathbf{\beta}_{old}+\mathbf{\Delta}_{old}^{-1}[\mathbf{y}-\mathbf{p}_{old}]. Here, on the penalized problem, we can easily prove that\frac{\partial\log\mathcal{L}_p(\mathbf{\beta}_{\lambda,old})}{\partial\mathbf{\beta}}=\frac{\partial\log\mathcal{L}(\mathbf{\beta}_{\lambda,old})}{\partial\mathbf{\beta}}-2\lambda\mathbf{\beta}_{old}while\frac{\partial^2\log\mathcal{L}_p(\mathbf{\beta}_{\lambda,old})}{\partial\mathbf{\beta}\partial\mathbf{\beta}^T}=\frac{\partial^2\log\mathcal{L}(\mathbf{\beta}_{\lambda,old})}{\partial\mathbf{\beta}\partial\mathbf{\beta}^T}-2\lambda\mathbb{I}Hence\mathbf{\beta}_{\lambda,new} =(\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{X}+2\lambda\mathbb{I})^{-1}\mathbf{X}^T\mathbf{\Delta}_{old}\mathbf{z}
The code is then

Y = myocarde$PRONO
X = myocarde[,1:7]
for(j in 1:7) X[,j] = (X[,j]-mean(X[,j]))/sd(X[,j])
X = as.matrix(X)
X = cbind(1,X)
colnames(X) = c("Inter",names(myocarde[,1:7]))
 beta = as.matrix(lm(Y~0+X)$coefficients,ncol=1)
 for(s in 1:9){
   pi = exp(X%*%beta[,s])/(1+exp(X%*%beta[,s]))
   Delta = matrix(0,nrow(X),nrow(X));diag(Delta)=(pi*(1-pi))
   z = X%*%beta[,s] + solve(Delta)%*%(Y-pi)
   B = solve(t(X)%*%Delta%*%X+2*lambda*diag(ncol(X))) %*% (t(X)%*%Delta%*%z)
   beta = cbind(beta,B)}
beta[,8:10]
              [,1]        [,2]        [,3]
XInter  0.59619654  0.59619654  0.59619654
XFRCAR  0.09217848  0.09217848  0.09217848
XINCAR  0.77165707  0.77165707  0.77165707
XINSYS  0.69678521  0.69678521  0.69678521
XPRDIA -0.29575642 -0.29575642 -0.29575642
XPAPUL -0.23921101 -0.23921101 -0.23921101
XPVENT -0.33120792 -0.33120792 -0.33120792
XREPUL -0.84308972 -0.84308972 -0.84308972

Again, it seems that convergence is very fast.

And interestingly, with that algorithm, we can also derive the variance of the estimator\text{Var}[\widehat{\mathbf{\beta}}_{\lambda}]=[\mathbf{X}^T\mathbf{\Delta}\mathbf{X}+2\lambda\mathbb{I}]^{-1}\mathbf{X}^T\mathbf{\Delta}\text{Var}[\mathbf{z}]\mathbf{\Delta}\mathbf{X}[\mathbf{X}^T\mathbf{\Delta}\mathbf{X}+2\lambda\mathbb{I}]^{-1}where\text{Var}[\mathbf{z}]=\mathbf{\Delta}^{-1}

The code to compute \widehat{\mathbf{\beta}}_{\lambda} as a function of \lambda is then

newton_ridge = function(lambda=1){
 beta = as.matrix(lm(Y~0+X)$coefficients,ncol=1)*runif(8)
 for(s in 1:20){
   pi = exp(X%*%beta[,s])/(1+exp(X%*%beta[,s]))
   Delta = matrix(0,nrow(X),nrow(X));diag(Delta)=(pi*(1-pi))
   z = X%*%beta[,s] + solve(Delta)%*%(Y-pi)
   B = solve(t(X)%*%Delta%*%X+2*lambda*diag(ncol(X))) %*% (t(X)%*%Delta%*%z)
   beta = cbind(beta,B)}
Varz = solve(Delta)
Varb = solve(t(X)%*%Delta%*%X+2*lambda*diag(ncol(X))) %*% t(X)%*% Delta %*% Varz %*%
  Delta %*% X %*% solve(t(X)%*%Delta%*%X+2*lambda*diag(ncol(X)))
return(list(beta=beta[,ncol(beta)],sd=sqrt(diag(Varb))))}

We can visualize the evolution of \widehat{\mathbf{\beta}}_{\lambda} (as a function of \lambda)

v_lambda=c(exp(seq(-2,5,length=61)))
est_ridge=Vectorize(function(x) newton_ridge(x)$beta)(v_lambda)
library("RColorBrewer")
colrs=brewer.pal(7,"Set1")
plot(v_lambda,est_ridge[1,],col=colrs[1],type="l")
for(i in 2:7) lines(v_lambda,est_ridge[i,],col=colrs[i])


and to get the evolution of the variance

v_lambda=c(exp(seq(-2,5,length=61)))
est_ridge=Vectorize(function(x) newton_ridge(x)$sd)(v_lambda)
library("RColorBrewer")
colrs=brewer.pal(7,"Set1")
plot(v_lambda,est_ridge[1,],col=colrs[1],type="l")
for(i in 2:7) lines(v_lambda,est_ridge[i,],col=colrs[i],lwd=2)


Recall that when \lambda=0 (on the left of the graphs), \widehat{\mathbf{\beta}}_{0}=\widehat{\mathbf{\beta}}^{mco} (no penalty). Thus as \lambda increase (i) the bias increase (estimates tend to 0) (ii) the variances deacrease.

Ridge, using glmnet

As always, there are R functions availble to run a ridge regression. Let us use the glmnet function, with \alpha=0

y = myocarde$PRONO
X = myocarde[,1:7]
for(j in 1:7) X[,j] = (X[,j]-mean(X[,j]))/sd(X[,j])
X = as.matrix(X)
library(glmnet)
glm_ridge = glmnet(X, y, alpha=0)
plot(glm_ridge,xvar="lambda",col=colrs,lwd=2)

as a function of the norm

the \ell_1 norm here, I don’t know why. I don’t know either why all graphs obtained with different optimisation routines are so different… Maybe that will be for another post…

Ridge with orthogonal covariates

An interesting case is obtained when covariates are orthogonal. This can be obtained using a PCA of the covariates.

library(factoextra)
pca = princomp(X)
pca_X = get_pca_ind(pca)$coord

Let us run a ridge regression on those (orthogonal) covariates

library(glmnet)
glm_ridge = glmnet(pca_X, y, alpha=0)
plot(glm_ridge,xvar="lambda",col=colrs,lwd=2)

plot(glm_ridge,col=colrs,lwd=2)

We clearly observe the shrinkage of the parameters, in the sense that \widehat{\mathbf{\beta}}_{\lambda}^{\perp}=\frac{\widehat{\mathbf{\beta}}^{mco}}{1+\lambda}

Application

Let us try with our second set of data

df0 = df
df0$y=as.numeric(df$y)-1
plot_lambda = function(lambda){
m = apply(df0,2,mean)
s = apply(df0,2,sd)
for(j in 1:2) df0[,j] = (df0[,j]-m[j])/s[j]
reg = glmnet(cbind(df0$x1,df0$x2), df0$y==1, alpha=0,lambda=lambda)
u = seq(0,1,length=101)
p = function(x,y){
  xt = (x-m[1])/s[1]
  yt = (y-m[2])/s[2]
  predict(reg,newx=cbind(x1=xt,x2=yt),type='response')}
v = outer(u,u,p)
image(u,u,v,col=clr10,breaks=(0:10)/10)
points(df$x1,df$x2,pch=c(1,19)[1+z],cex=1.5)
contour(u,u,v,levels = .5,add=TRUE)
}

We can try various values of \lambda

reg = glmnet(cbind(df0$x1,df0$x2), df0$y==1, alpha=0)
par(mfrow=c(1,2))
plot(reg,xvar="lambda",col=c("blue","red"),lwd=2)
abline(v=log(.2))
plot_lambda(.2)


or

reg = glmnet(cbind(df0$x1,df0$x2), df0$y==1, alpha=0)
par(mfrow=c(1,2))
plot(reg,xvar="lambda",col=c("blue","red"),lwd=2)
abline(v=log(1.2))
plot_lambda(1.2)


Next step is to change the norm of the penality, with the \ell_1 norm (to be continued…)

Enveloppe convexe de points tirés au hasard

Le week-end dernier, Jean-Baptiste qui était de passage à la maison, me présentait un problème amusant de géométrie, lié à un papier mis en ligne l’an dernier, monotonicity of facet numbers of random convex hulls. Dans cet article, ils montrent que quand on tire n points au hasard (dans un espace de dimension d) alors P_n, le nombre moyen de face de l’enveloppe convexe est strictement croissant avec n. Si on n’a pas regardé la démonstration (on avait mieux à faire), Jean-Baptiste me disait que ce problème très simple était en fait très complexe. Et bien entendu, comme ça m’a interpelé, j’ai voulu regarder plus en détails. En dimension d=2, et en tirant en plus uniformément sur le carré unité (oui, j’ai fait très très simple).

Pour tirer des points au hasard, et récupérer l’enveloppe convexe, c’est assez simple, par exemple avec 15 points

library(sp)
library(geosphere)
n=15
UV=matrix(runif(2*n),n,2)
CH=chull(UV)
PLCH=UV[c(CH,CH[1]),]
plot(c(0,0,1,1),c(1,0,0,1))
polygon(PLCH,border="blue",col=rgb(0,0,1,.3))
points(UV,pch=19,cex=2,col="red")

On le voit, ici l’enveloppe convexe est un polygône de 8 côtés (ou 8 sommets). On peut alors faire un petit code pour tirer des points au hasard, construire l’enveloppe convexe, et sortir des infos (nombre de points extrémaux, surface de l’enveloppe convexe, présence ou non de certains points – sur la diagonale, etc)

simu=function(n,isplot=FALSE){
UV=matrix(runif(2*n),n,2)
CH=chull(UV)
PLCH=UV[c(CH,CH[1]),]
nb_ex=length(CH)
p_in=function(u) point.in.polygon(u,u,PLCH[,1],PLCH[,2])
pts_in=Vectorize(p_in)(seq(.5,.95,by=.05))
if(isplot==TRUE) lines(PLCH,col=rgb(0,0,1,.25))
return(list(nb=nb_ex,area=areaPolygon(PLCH),pts=pts_in))}

par exemple

plot(c(0,0,1,1),c(1,0,0,1),col="white")
for(s in 1:1000){S=simu(5,isplot=TRUE)}

(on pourrait bien entendu stocker tout plein de choses)

ou encore avec n= 20 points au lieu de 5

plot(c(0,0,1,1),c(1,0,0,1),col="white")
for(s in 1:1000){S=simu(20,isplot=TRUE)}

Essayons de boucler un peu sur n maintenant

Np=c(3,4,5,6,7,8,10,15,20,30,40,50,75,100,200)
VN=VA=rep(NA,15)
NN=matrix(NA,20000,15)
VPT=matrix(NA,15,10)
for(i in 1:15){
N=A=rep(NA,20000)
PT=matrix(NA,20000,10)
np=Np[i]
for(s in 1:20000){
S=simu(np,isplot=FALSE)
N[s]=S$nb
PT[s,]=S$pts
A[s]=S$area
}
NN[,i]=N
VN[i]=mean(N,na.rm=TRUE)
VA[i]=mean(A,na.rm=TRUE)
VPT[i,]=apply(PT,2,function(x) mean(x,na.rm=TRUE))
}

Cette fois on stocke tout plein de choses. On peut juste faire un boxplot du nombre des points extrémaux en fonction de la taille de l’échantillon

VV=rep(Np,each=20000)
boxplot(as.vector(NN)~as.factor(VV))

Oui, en moyenne, ça semble croitre. Plus amusant, si on regarde la moyenne en fonction de \log(n)

plot(Np,VN,type="l",log="x",col="blue")

on obtient… une belle droite ! Le nombre moyen de points extrémaux croit en \log(n). On peut même avoir la pente de cette droite,

> lm(VN~log(Np))
Coefficients:
(Intercept) log(Np)
0.05224 2.58717717

Je laisse les plus courageux trouver du sens à ce 2.58717… Si on continue un peu, on peut regarder la probabilité que (u,u) soit à l’intérieur, pour plusieurs valeurs de u.

plot(Np,VPT[,10],type="l",log="x",col="blue")
lines(Np,VPT[,8])
lines(Np,VPT[,5],col="red",lwd=2)

On retrouve des fonctions croisantes, en fonction de n, mais la convexité semble dépendre de l’endroit où se trouve le point u. Amusant, non ?

Nonconvexity, and playing indoor paintball

Following the two previous posts (here and there), on the number of people that don’t get wet while playing with water pistols, consider now an indoor version, in a non-convex room (i.e. player behind wall are now, somehow, protected). In the previous posts, players where playing on a square field, and I briefly mentioned that if the field was a disk, results would have been (roughly) the same: so far, the shape of the field was not an issue. But what if the field is no longer convex,

library(sp)
plot(0:2,0:2,col="white",xlab="",ylab="")
MAP=Polygon(cbind(c(0,0,1,1,2,2,0),
c(0,2,2,1,1,0,0)))
polygon(MAP@coords,col="light blue")

and players hidden behind the wall cannot be reached (red lines above are impossible hits). As earlier, it is still possible to look at the closest neighbor, we just have to exclude pairs that can no longer hit each other.

And again, it is possible to plot safe zones in green.

Once again, it is possible to look more closely are those supposed-to-be “safe zones”, i.e. by looking at the distribution of the location of players that were dry at the end of the game. With 11 players, we obtain


What about the distribution of the number of dry players, over a game ?

touch=function(x1,y1,x2,y2,n=251){
X=seq(x1,x2,length=n)
Y=seq(y1,y2,length=n)
sum(point.in.polygon(X,Y,MAP@coords[,1],
MAP@coords[,2], mode.checked=FALSE)==0)==0
}

NOTWETnc=function(n,p){
sx=runif(50)*2;sy=runif(50)*2
IN=which(point.in.polygon(sx,sy,MAP@coords[,1],
MAP@coords[,2], mode.checked=FALSE)==1)
Sx=sx[IN];Sy=sy[IN]
Sx=Sx[1:n];Sy=Sy[1:n]
IN=IN[1:n]
MI=matrix(NA,n,n)
for(i in 1:(n-1)){
for(j in (i+1):(n)){
MI[j,i]=MI[i,j]=touch(Sx[i],Sy[i],Sx[j],Sy[j])
}}
(d=as.matrix(dist(cbind(Sx,Sy),
method = "euclidean",upper=TRUE)))
diag(d)=999999
dpossible=d
dpossible[MI==FALSE]=999999
dmin=apply(dpossible,2,which.min)
#whonotwet=( (1:n) %notin% names(table(dmin)) )
notwet=n-length(table(dmin))
return(notwet)}

NOTWET=function(n){
x=runif(n)
y=runif(n)
(d=as.matrix(dist(cbind(x,y),
method = "euclidean",upper=TRUE)))
diag(d)=999999
dmin=apply(d,2,which.min)
notwet=n-length(table(dmin))
return(notwet)}

NSim=10000
Nnc=Vectorize(NOTWETnc)(n=rep(11,NSim))
Nc=Vectorize(NOTWET)(n=rep(11,NSim))
T=table(Nc)
Tn=table(Nnc)
plot(as.numeric(names(Tn)),
Tn/NSim,type="b",col="blue")
lines(as.numeric(names(T)),
T/NSim,type="b",col="red",pch=4)

On 11 players, we have the same distribution as the one on a square field. So convexity is not a key issue here…

Strange isn’t it. And with an odd number of player, not only there is at least one dry player, but at least, half of the players (maybe minus one) have to be wet…