## 2.4 Bandwidth selection

As we saw in the previous sections, the kde critically depends on the employed bandwidth. The purpose of this section is to introduce objective and automatic bandwidth selectors that attempt to minimize the estimation error of the target density \(f\).

The first step is to define a global, rather than local, error criterion. The *Integrated Squared Error* (ISE),

is the squared distance between the kde and the target density. The ISE is a random quantity, since it depends directly on the sample \(X_1,\ldots,X_n\). As a consequence, looking for an optimal-ISE bandwidth is a hard task, since the optimality is dependent on the sample itself and not only on the population and \(n\). To avoid this problem, it is usual to compute the *Mean Integrated Squared Error* (MISE):

The MISE is convenient due to its mathematical tractability and its natural relation with the MSE. There are, however, other error criteria that present attractive properties, such as the *Mean Integrated Absolute Error* (MIAE):

The MIAE, unlike the MISE, is *invariant* with respect to monotone transformations of the data. For example, if \(g(x)=f(t^{-1}(x))(t^{-1})'(x)\) is the density of \(Y=t(X)\) and \(X\sim f\), then if the change of variables \(y=t(x)\) is made^{31},

Despite this appealing property, the analysis of MIAE is substantially more complicated than the MISE. We refer to Devroye and Györfi (1985) for a comprehensive treatment of absolute value metrics for kde.

Once the MISE is set as the error criterion to be minimized, our aim is to find

\[\begin{align*} h_\mathrm{MISE}:=\arg\min_{h>0}\mathrm{MISE}[\hat{f}(\cdot;h)]. \end{align*}\]For that purpose, we need an explicit expression of the MISE that we can attempt to minimize. The next asymptotic expansion for the MISE solves this issue.

**Corollary 2.2 **Under **A1**–**A3**,

Therefore, \(\mathrm{MISE}[\hat{f}(\cdot;h)]\to0\) when \(n\to\infty\).

**Exercise 2.11 **Prove the corollary by integrating the MSE of \(\hat{f}(x;h)\).

The dominating part of the MISE is denoted as AMISE, which stands for *Asymptotic MISE*:

Thanks to the closed expression of the AMISE, it is possible to obtain a bandwidth that minimizes this error.

**Corollary 2.3 **The bandwidth that minimizes the AMISE is

The optimal AMISE is:

\[\begin{align} \inf_{h>0}\mathrm{AMISE}[\hat{f}(\cdot;h)]=\frac{5}{4}(\mu_2^2(K)R(K)^4)^{4/5}R(f'')^{1/5}n^{-4/5}.\tag{2.23} \end{align}\]**Exercise 2.12 **Prove the corollary by differentiating \(\mathrm{AMISE}[\hat{f}(\cdot;h)]\) with respect to \(h\).

The AMISE-optimal order deserves some further inspection. It can be seen in Section 3.2 of Scott (2015) that the AMISE-optimal order for the *histogram* of Section 2.1 is \(\left(\frac{3}{4}\right)^{2/3}R(f')^{1/3}n^{-2/3}\). Two aspects are of interest when comparing this result with (2.23):

- First, the MISE of the histogram is asymptotically larger than the MISE of the kde
^{32}. This is a quantification of the quite apparent visual improvement of the kde over the histogram. - Second, \(R(f')\) appears instead of \(R(f'')\), evidencing that the histogram is affected not only by the
*curvature*of the target density \(f\) but also by*how fast \(f\) varies*.

Unfortunately, the AMISE bandwidth depends on \(R(f'')=\int(f''(x))^2\,\mathrm{d}x\), which measures the *curvature* of the density. As a consequence, it can not be readily applied in practice, as \(R(f'')\) is unknown! In the next subsection we will see how to plug-in estimates for \(R(f'')\).

### 2.4.1 Plug-in rules

A simple solution to estimate \(R(f'')\) is to assume that \(f\) is the density of a \(\mathcal{N}(\mu,\sigma^2)\), and then plug-in the form of the curvature for such density:

\[\begin{align*} R(\phi''_\sigma(\cdot-\mu))=\frac{3}{8\pi^{1/2}\sigma^5}. \end{align*}\]While doing so, we approximate the curvature of an arbitrary density by means of the curvature of a Normal and we have that

\[\begin{align*} h_\mathrm{AMISE}=\left[\frac{8\pi^{1/2}R(K)}{3\mu_2^2(K)n}\right]^{1/5}\sigma. \end{align*}\]Interestingly, the bandwidth is directly proportional to the standard deviation of the target density. Replacing \(\sigma\) by an estimate yields the **normal scale bandwidth selector**, which we denote by \(\hat{h}_\mathrm{NS}\) to emphasize its randomness:

The estimate \(\hat\sigma\) can be chosen as the standard deviation \(s\), or, in order to avoid the effects of potential outliers, as the standardized interquantile range

\[\begin{align*} \hat \sigma_{\mathrm{IQR}}:=\frac{X_{([0.75n])}-X_{([0.25n])}}{\Phi^{-1}(0.75)-\Phi^{-1}(0.25)} \end{align*}\]or as

\[\begin{align} \hat\sigma=\min(s,\hat \sigma_{\mathrm{IQR}}). \tag{2.24} \end{align}\]When combined with a normal kernel, for which \(\mu_2(K)=1\) and \(R(K)=\frac{1}{2\sqrt{\pi}}\), this particularization of \(\hat{h}_{\mathrm{NS}}\) gives the famous **rule-of-thumb** for bandwidth selection:

\(\hat{h}_{\mathrm{RT}}\) is implemented in R through the function `bw.nrd`

^{33}.

```
# Data
set.seed(667478)
n <- 100
x <- rnorm(n)
# Rule-of-thumb
bw.nrd(x = x)
## [1] 0.4040319
# bwd.nrd employs 1.34 as an approximation for diff(qnorm(c(0.25, 0.75)))
# Same as
iqr <- diff(quantile(x, c(0.25, 0.75))) / diff(qnorm(c(0.25, 0.75)))
1.06 * n^(-1/5) * min(sd(x), iqr)
## [1] 0.4040319
```

The previous selector is an example of a **zero-stage plug-in** selector, a terminology inspired by the fact that \(R(f'')\) was estimated by plugging-in a parametric assumption at the “very beginning”. Another possibility could have been to estimate \(R(f'')\) nonparametrically and then plug-in the estimate into into \(h_\mathrm{AMISE}\). Let’s explore this possibility in more detail next. But first, note the useful equality

This holds by a iterative application of integration by parts. For example, for \(s=2\), take \(u=f''(x)\) and \(\,\mathrm{d}v=f''(x)\,\mathrm{d}x\). It gives

\[\begin{align*} \int f''(x)^2\,\mathrm{d}x&=[f''(x)f'(x)]_{-\infty}^{+\infty}-\int f'(x)f'''(x)\,\mathrm{d}x\\ &=-\int f'(x)f'''(x)\,\mathrm{d}x \end{align*}\]under the assumption that the derivatives vanish at infinity. Applying again integration by parts with \(u=f'''(x)\) and \(\,\mathrm{d}v=f'(x)\,\mathrm{d}x\) gives the result. This simple derivation has an important consequence: for estimating the functionals \(R(f^{(s)})\) it suffices to estimate for \(r=2s\) the functionals

\[\begin{align*} \psi_r:=\int f^{(r)}(x)f(x)\,\mathrm{d}x=\mathbb{E}[f^{(r)}(X)]. \end{align*}\]Particularly, \(R(f'')=\psi_4\).

Thanks to the previous expression, a possible way to estimate \(\psi_r\) nonparametrically is

\[\begin{align} \hat\psi_r(g)&=\frac{1}{n}\sum_{i=1}^n\hat{f}^{(r)}(X_i;g)\nonumber\\ &=\frac{1}{n^2}\sum_{i=1}^n\sum_{j=1}^nL_g^{(r)}(X_i-X_j),\tag{2.25} \end{align}\]where \(\hat{f}^{(r)}(\cdot;g)\) is the \(r\)-th derivative of a kde with bandwidth \(g\) and kernel \(L\), *i.e.*

Note that \(g\) and \(L\) can be different from \(h\) and \(K\), respectively. It turns out that estimating \(\psi_r\) involves the adequate selection of a bandwidth \(g\). The agenda is analogous to the one for \(h_\mathrm{AMISE}\), but now taking into account that both \(\hat\psi_r(g)\) and \(\psi_r\) are *scalar* quantities:

Under certain regularity assumptions

\[\begin{align*} \mathrm{AMSE}[\hat \psi_r(g)]=&\,\left\{\frac{L^{(r)}(0)}{ng^{r+1}}+\frac{\mu_2(L)\psi_{r+2}g^2}{4}\right\}+\frac{2R(L^{(r)})\psi_0}{n^2g^{2r+1}}\\ &+\frac{4}{n}\left\{\int f^{(r)}(x)^2f(x)\,\mathrm{d}x-\psi_r^2\right\}. \end{align*}\]^{34}, the asymptotic bias and variance of \(\hat\psi_r(g)\) are obtained. With them, we can compute the asymptotic expansion of the MSE and obtain the*Asymptotic Mean Squared Error*AMSE:*Note*: \(k\) is the highest integer such that \(\mu_k(L)>0\). In these notes we have restricted to the case \(k=2\) for the kernels \(K\), but there are theoretical gains if one allows*high-order kernels*\(L\) with vanishing even moments larger than \(2\) for estimating \(\psi_r\).Obtain the AMSE-optimal bandwidth:

\[\begin{align*} g_\mathrm{AMSE}=\left[-\frac{k!L^{(r)}(0)}{\mu_k(L)\psi_{r+k}n}\right]^{1/(r+k+1)} \end{align*}\]The order of the optimal AMSE is

\[\begin{align*}\inf_{g>0}\mathrm{AMSE}[\hat \psi_r(g)]=\begin{cases} O(n^{-(2k+1)/(r+k+1)}),&k<r,\\ O(n^{-1}),&k\geq r, \end{cases}\end{align*}\]which shows that a parametric-like rate of convergence can be achieved with high-order kernels. If we consider \(L=K\) and \(k=2\), then

\[\begin{align*} g_\mathrm{AMSE}=\left[-\frac{2K^{(r)}(0)}{\mu_2(L)\psi_{r+2}n}\right]^{1/(r+3)}. \end{align*}\]

The above result has a major problem: it depends on \(\psi_{r+2}\)! Thus if we want to estimate \(R(f'')=\psi_4\) by \(\hat\psi_4(g_\mathrm{AMSE})\) we will need to estimate \(\psi_6\). But \(\hat\psi_6(g_\mathrm{AMSE})\) will depend on \(\psi_8\) and so on. The solution to this convoluted problem is to stop estimating the functional \(\psi_r\) after a given number \(\ell\) of *stages*, therefore the terminology **\(\ell\)-stage plug-in selector**. At the \(\ell\) stage, the functional \(\psi_{\ell+4}\) in the AMSE-optimal bandwidth for estimating \(\psi_{\ell+2}\) is computed assuming that the density is a \(\mathcal{N}(\mu,\sigma^2)\), for which

Typically, **two stages** are considered a good trade-off between bias (mitigated when \(\ell\) increases) and variance (increases with \(\ell\)) of the plug-in selector. This is the method proposed by Sheather and Jones (1991), where they consider \(L=K\) and \(k=2\), yielding what we call the **direct plug-in** (DPI). The algorithm is:

- Estimate \(\psi_8\) using \(\hat\psi_8^\mathrm{NS}:=\frac{105}{32\sqrt{\pi}\hat\sigma^9}\), where \(\hat\sigma\) is given in (2.24)
Estimate \(\psi_6\) using \(\hat\psi_6(g_1)\) from (2.25), where

\[\begin{align*} g_1:=\left[-\frac{2K^{(6)}(0)}{\mu_2(K)\hat\psi^\mathrm{NS}_{8}n}\right]^{1/9}. \end{align*}\]Estimate \(\psi_4\) using \(\hat\psi_4(g_2)\) from (2.25), where

\[\begin{align*} g_2:=\left[-\frac{2K^{(4)}(0)}{\mu_2(K)\hat\psi_6(g_1)n}\right]^{1/7}. \end{align*}\]The selected bandwidth is

\[\begin{align*} \hat{h}_{\mathrm{DPI}}:=\left[\frac{R(K)}{\mu_2^2(K)\hat\psi_4(g_2)n}\right]^{1/5}. \end{align*}\]

*Remark. * The derivatives \(K^{(r)}\) for the normal kernel can be obtained using *Hermite polynomials*: \(\phi^{(r)}(x)=\phi(x)H_r(x)\). For \(r=4,6\), \(H_4(x)=x^4-6x^2+3\) and \(H_6(x)=x^6-16x^4+45x^2-15\).

\(\hat{h}_{\mathrm{DPI}}\) is implemented in R through the function `bw.SJ`

(use `method = "dpi"`

). An alternative and faster implementation is `ks::hpi`

, which also for more flexibility and has a somehow more complete documentation.

```
# Data
set.seed(672641)
x <- rnorm(100)
# DPI selector
bw.SJ(x = x, method = "dpi")
## [1] 0.5006905
# Similar to
ks::hpi(x) # Default is two-stages
## [1] 0.4999456
```

**Exercise 2.13 **Apply and inspect the kde of `airquality$Ozone`

using the DPI selector (both `bw.SJ`

and `ks::hpi`

). What can you conclude from the estimate? *Beware*: remove `NA`

s prior to performing the kde and bandwidth selection.

### 2.4.2 Cross-validation

We turn now our attention to a different philosophy of bandwidth estimation. Instead of trying to minimize the AMISE by plugging-in estimates for the unknown curvature term, we directly attempt to minimize the MISE. The idea is to use the sample twice: one for computing the kde and other for *evaluating* its performance on estimating \(f\). To avoid the clear dependence on the sample, we do this evaluation in a *cross-validatory* way: the data used for computing the kde is *not* used for its evaluation.

We begin by expanding the square in the MISE expression:

\[\begin{align*} \mathrm{MISE}[\hat{f}(\cdot;h)]=&\,\mathbb{E}\left[\int (\hat{f}(x;h)-f(x))^2\,\mathrm{d}x\right]\\ =&\,\mathbb{E}\left[\int \hat{f}(x;h)^2\,\mathrm{d}x\right]-2\mathbb{E}\left[\int \hat{f}(x;h)f(x)\,\mathrm{d}x\right]\\ &+\int f(x)^2\,\mathrm{d}x. \end{align*}\]Since the last term does not depend on \(h\), minimizing \(\mathrm{MISE}[\hat{f}(\cdot;h)]\) is equivalent to minimizing

\[\begin{align*} \mathbb{E}\left[\int \hat{f}(x;h)^2\,\mathrm{d}x\right]-2\mathbb{E}\left[\int \hat{f}(x;h)f(x)\,\mathrm{d}x\right]. \end{align*}\]This quantity is unknown, but it can be estimated unbiasedly by

\[\begin{align} \mathrm{LSCV}(h):=\int\hat{f}(x;h)^2\,\mathrm{d}x-2n^{-1}\sum_{i=1}^n\hat{f}_{-i}(X_i;h),\tag{2.26} \end{align}\]where \(\hat{f}_{-i}(\cdot;h)\) is the *leave-one-out* kde and is based on the sample with the \(X_i\) removed:

The motivation for (2.26) is the following. The first term is unbiased by design^{35}. The second arises from approximating \(\int \hat{f}(x;h)f(x)\,\mathrm{d}x\) by Monte Carlo from the sample \(X_1,\ldots,X_n\sim f\), or in other words, by replacing \(f(x)\,\mathrm{d}x=\,\mathrm{d}F(x)\) with \(\mathrm{d}F_n(x)\). This gives

and, in order to mitigate the dependence of the sample, we replace \(\hat{f}(X_i;h)\) by \(\hat{f}_{-i}(X_i;h)\) above. In that way, we use the sample for estimating the integral involving \(\hat{f}(\cdot;h)\), but for each \(X_i\) we compute the kde on the *remaining* points.

The **least squares cross-validation** (LSCV) selector, also denoted **unbiased cross-validation** (UCV) selector, is defined as

Numerical optimization is required for obtaining \(\hat{h}_\mathrm{LSCV}\), contrary to the previous plug-in selectors, and there is little control on the shape of the objective function. This will be also the case for the forthcoming bandwidth selectors. The following remark warns about the dangers of numerical optimization in this context.

*Remark. * Numerical optimization of the LSCV function can be challenging. In practice, several local minima are possible, and the roughness of the objective function can vary notably depending on \(n\) and \(f\). As a consequence, optimization routines may get trapped in spurious solutions. To be on the safe side, it is always advisable to check (when possible) the solution by plotting \(\mathrm{LSCV}(h)\) for a range of \(h\), or to perform a search in a bandwidth grid: \(\hat{h}_\mathrm{LSCV}\approx\arg\min_{h_1,\ldots,h_G}\mathrm{LSCV}(h)\).

\(\hat{h}_{\mathrm{LSCV}}\) is implemented in R through the function `bw.ucv`

. `bw.ucv`

uses R’s `optimize`

, which is quite sensible to the selection of the search interval (long intervals containing the solution may lead to unsatisfactory termination of the search; and short intervals might not contain the minimum). Therefore, some care is needed and that is why the `bw.ucv.mod`

function is presented below.

```
# Data
set.seed(123456)
x <- rnorm(100)
# UCV gives a warning
bw.ucv(x = x)
## [1] 0.4499177
# Extend search interval
bw.ucv(x = x, lower = 0.01, upper = 1)
## [1] 0.5482419
# bw.ucv.mod replaces the optimization routine of bw.ucv by an exhaustive
# search on "h.grid" (chosen adaptatively from the sample) and optionally
# plots the LSCV curve with "plot.cv"
bw.ucv.mod <- function(x, nb = 1000L,
h_grid = diff(range(x)) * (seq(0.1, 1, l = 200))^2,
plot_cv = FALSE) {
if ((n <- length(x)) < 2L)
stop("need at least 2 data points")
n <- as.integer(n)
if (is.na(n))
stop("invalid length(x)")
if (!is.numeric(x))
stop("invalid 'x'")
nb <- as.integer(nb)
if (is.na(nb) || nb <= 0L)
stop("invalid 'nb'")
storage.mode(x) <- "double"
hmax <- 1.144 * sqrt(var(x)) * n^(-1/5)
Z <- .Call(stats:::C_bw_den, nb, x)
d <- Z[[1L]]
cnt <- Z[[2L]]
fucv <- function(h) .Call(stats:::C_bw_ucv, n, d, cnt, h)
## Original
# h <- optimize(fucv, c(lower, upper), tol = tol)$minimum
# if (h < lower + tol | h > upper - tol)
# warning("minimum occurred at one end of the range")
## Modification
obj <- sapply(h_grid, function(h) fucv(h))
h <- h_grid[which.min(obj)]
if (plot_cv) {
plot(h_grid, obj, type = "o")
rug(h_grid)
abline(v = h, col = 2, lwd = 2)
}
h
}
# Compute the bandwidth and plot the LSCV curve
bw.ucv.mod(x = x, plot_cv = TRUE)
## [1] 0.5431732
# We can compare with the default bw.ucv output
abline(v = bw.ucv(x = x), col = 3)
```

The next cross-validation selector is based on **biased cross-validation** (BCV). The BCV selector presents a *hybrid strategy* that combines plug-in and cross-validation ideas. It starts by considering the AMISE expression in (2.22)

and then plugs-in an estimate for \(R(f'')\) based on a modification of \(R(\hat{f}\!'' (\cdot;h))\). The modification is

\[\begin{align} \widetilde{R(f'')}:=&\,R(\hat{f}\!''(\cdot;h))-\frac{R(K'')}{nh^5}\nonumber\\ =&\,\frac{1}{n}\sum_{i=1}^n\sum_{\substack{j=1\\j\neq i}}^n(K_h''*K_h'')(X_i-X_j)\tag{2.27}, \end{align}\]a *leave-out-diagonals* estimate of \(R(f'')\). It is designed to reduce the bias of \(R(\hat{f}''(\cdot;h))\), since \(\mathbb{E}\left[R(\hat{f}''(\cdot;h))\right]=R(f'')+\frac{R(K'')}{nh^5}+O(h^2)\) (Scott and Terrell 1987). Plugging-in (2.27) into the AMISE expression yields the BCV objective function and the BCV bandwidth selector:

The appealing property of \(\hat{h}_\mathrm{BCV}\) is that it has a considerably smaller variance compared to \(\hat{h}_\mathrm{LSCV}\). This reduction in variance comes at the price of an increased bias, which tends to make \(\hat{h}_\mathrm{BCV}\) larger than \(h_\mathrm{MISE}\).

\(\hat{h}_{\mathrm{BCV}}\) is implemented in R through the function `bw.bcv`

. Again, `bw.bcv`

uses R’s `optimize`

so the `bw.bcv.mod`

function is presented to have better guarantees on finding the adequate minimum.

```
# Data
set.seed(123456)
x <- rnorm(100)
# BCV gives a warning
bw.bcv(x = x)
## [1] 0.4500924
# Extend search interval
args(bw.bcv)
## function (x, nb = 1000L, lower = 0.1 * hmax, upper = hmax, tol = 0.1 *
## lower)
## NULL
bw.bcv(x = x, lower = 0.01, upper = 1)
## [1] 0.5070129
# bw.bcv.mod replaces the optimization routine of bw.bcv by an exhaustive
# search on "h.grid" (chosen adaptatively from the sample) and optionally
# plots the BCV curve with "plot.cv"
bw.bcv.mod <- function(x, nb = 1000L,
h_grid = diff(range(x)) * (seq(0.1, 1, l = 200))^2,
plot_cv = FALSE) {
if ((n <- length(x)) < 2L)
stop("need at least 2 data points")
n <- as.integer(n)
if (is.na(n))
stop("invalid length(x)")
if (!is.numeric(x))
stop("invalid 'x'")
nb <- as.integer(nb)
if (is.na(nb) || nb <= 0L)
stop("invalid 'nb'")
storage.mode(x) <- "double"
hmax <- 1.144 * sqrt(var(x)) * n^(-1/5)
Z <- .Call(stats:::C_bw_den, nb, x)
d <- Z[[1L]]
cnt <- Z[[2L]]
fbcv <- function(h) .Call(stats:::C_bw_bcv, n, d, cnt, h)
## Original code
# h <- optimize(fbcv, c(lower, upper), tol = tol)$minimum
# if (h < lower + tol | h > upper - tol)
# warning("minimum occurred at one end of the range")
## Modification
obj <- sapply(h_grid, function(h) fbcv(h))
h <- h_grid[which.min(obj)]
if (plot_cv) {
plot(h_grid, obj, type = "o")
rug(h_grid)
abline(v = h, col = 2, lwd = 2)
}
h
}
# Compute the bandwidth and plot the BCV curve
bw.bcv.mod(x = x, plot_cv = TRUE)
## [1] 0.5130493
# We can compare with the default bw.bcv output
abline(v = bw.bcv(x = x), col = 3)
```

### 2.4.3 Comparison of bandwidth selectors

We state next some insights from the convergence results of the DPI, LSCV, and BCV selectors. All of them are based in results of the kind

\[\begin{align} n^\nu(\hat{h}/h_\mathrm{MISE}-1)\stackrel{d}{\longrightarrow}\mathcal{N}(0,\sigma^2),\tag{2.28} \end{align}\]where \(\sigma^2\) depends only on \(K\) and \(f\), and measures how variable is the selector. The rate \(n^\nu\) serves to quantify how fast^{36} the relative error \(\hat{h}/h_\mathrm{MISE}-1\) decreases (the larger the \(\nu\), the faster the convergence).

Under certain regularity conditions, we have:

\(n^{1/10}(\hat{h}_\mathrm{LSCV}/h_\mathrm{MISE}-1)\stackrel{d}{\longrightarrow}\mathcal{N}(0,\sigma_\mathrm{LSCV}^2)\) and \(n^{1/10}(\hat{h}_\mathrm{BCV}/h_\mathrm{MISE}-1)\stackrel{d}{\longrightarrow}\mathcal{N}(0,\sigma_\mathrm{BCV}^2)\). Both cross-validation selectors have a slow rate of convergence (compare it with the \(n^{1/2}\) of the CLT). Inspection of the variances of both selectors reveals that, for the normal kernel \(\sigma_\mathrm{LSCV}^2/\sigma_\mathrm{BCV}^2\approx 15.7\). Therefore,

**LSCV is considerably more variable than BCV**.\(n^{5/14}(\hat{h}_\mathrm{DPI}/h_\mathrm{MISE}-1)\stackrel{d}{\longrightarrow}\mathcal{N}(0,\sigma_\mathrm{DPI}^2)\). Thus, the

\[\begin{align*} \frac{1}{4}\mu_2^2(K)\tilde\psi_4(h)h^4+\frac{R(K)}{nh} \end{align*}\]**DPI selector**has a**convergence rate much faster**than the**cross-validation selectors**. There is an appealing explanation for this phenomenon. Recall that \(\hat{h}_\mathrm{BCV}\) minimizes the slightly modified version of \(\mathrm{BCV}(h)\) given byand

\[\begin{align*} \tilde\psi_4(h):=&\frac{1}{n(n-1)}\sum_{i=1}^n\sum_{\substack{j=1\\j\neq i}}^n(K_h''*K_h'')(X_i-X_j)\nonumber\\ =&\frac{n}{n-1}\widetilde{R(f'')}. \end{align*}\]\(\tilde\psi_4\) is a leave-out-diagonals estimate of \(\psi_4\). Despite being different from \(\hat\psi_4\), it serves for building a DPI analogous to BCV that points towards the precise fact that drags down the performance of BCV. The modified version of the DPI minimizes

\[\begin{align*} \frac{1}{4}\mu_2^2(K)\tilde\psi_4(g)h^4+\frac{R(K)}{nh}, \end{align*}\]where \(g\) is independent of \(h\). The two methods differ on the the way \(g\) is chosen: BCV sets \(g=h\) and the modified DPI looks for the best \(g\) in terms of the \(\mathrm{AMSE}[\tilde\psi_4(g)]\). It can be seen that \(g_\mathrm{AMSE}=O(n^{-2/13})\), whereas the \(h\) used in BCV is asymptotically \(O(n^{-1/5})\). This suboptimality on the choice of \(g\) is the reason of the asymptotic deficiency of BCV.

We focus now on exploring the empirical performance of bandwidth selectors. The workhorse for doing that is simulation. A popular collection of simulation scenarios was given by Marron and Wand (1992) and are conveniently available through the package `nor1mix`

. They form a collection of normal \(r\)-mixtures of the form

where \(w_j\geq0\), \(j=1,\ldots,r\) and \(\sum_{j=1}^rw_j=1\). Densities of this form are specially attractive since they allow for arbitrarily flexibility and, if the normal kernel is employed, they allow for *explicit and exact* MISE expressions:

These exact MISE expressions are highly convenient for computing \(h_\mathrm{MISE}\) easily by optimization of (2.29).

```
# Available models
?nor1mix::MarronWand
# Simulating
samp <- nor1mix::rnorMix(n = 500, obj = nor1mix::MW.nm9)
# MW object in the second argument
hist(samp, freq = FALSE)
# Density evaluation
x <- seq(-4, 4, length.out = 400)
lines(x, nor1mix::dnorMix(x = x, obj = nor1mix::MW.nm9), col = 2)
```

```
# Plot a MW object directly
# A normal with the same mean and variance is plotted in dashed lines
par(mfrow = c(2, 2))
plot(nor1mix::MW.nm5)
plot(nor1mix::MW.nm7)
plot(nor1mix::MW.nm10)
plot(nor1mix::MW.nm12)
lines(nor1mix::MW.nm10) # Also possible
```

**Exercise 2.14 **Implement the \(h_\mathrm{MISE}\) using (2.29) for model `nor1mix::MW.nm5`

. Then, investigate by simulation the distributions of \(\hat{h}_\mathrm{DPI}/h_\mathrm{MISE}-1\), \(\hat{h}_\mathrm{LSCV}/h_\mathrm{MISE}-1\), and \(\hat{h}_\mathrm{BCV}/h_\mathrm{MISE}-1\).

Figure 2.9 presents a visualization of the performance of the kde with different bandwidth selectors, carried out in the family of mixtures of Marron and Wand (1992).

**Which bandwidth selector is the most adequate for a given dataset?**

There is no simple and universal answer to this question. There are, however, a series of useful facts and suggestions:

- Trying several selectors and inspecting the results may help on determining which one is estimating the density better.
- The
**DPI selector has a convergence rate much faster than the cross-validation selectors**. Therefore, in theory is expected to perform better than LSCV and BCV. For this reason, it tends to be amongst the**preferred bandwidth selectors**in the literature. - Cross-validatory selectors may be better suited for highly non-normal and rough densities, in which plug-in selectors may end up oversmoothing.
- LSCV tends to be considerably more variable than BCV.
- The RT is a quick, simple, and inexpensive selector. However, it tends to give bandwidths that are too large for non-normal data.

### References

Devroye, L., and L. Györfi. 1985. *Nonparametric Density Estimation*. Wiley Series in Probability and Mathematical Statistics: Tracts on Probability and Statistics. New York: John Wiley & Sons, Inc.

Scott, D. W. 2015. *Multivariate Density Estimation*. Second. Wiley Series in Probability and Statistics. Hoboken: John Wiley & Sons, Inc. doi:10.1002/9781118575574.

Sheather, S. J., and M. C. Jones. 1991. “A Reliable Data-Based Bandwidth Selection Method for Kernel Density Estimation.” *J. Roy. Statist. Soc. Ser. B* 53 (3): 683–90. doi:10.1111/j.2517-6161.1991.tb01857.x.

Scott, D. W., and G. R. Terrell. 1987. “Biased and Unbiased Cross-Validation in Density Estimation.” *J. Am. Stat. Assoc.* 82 (400): 1131–46. doi:10.1080/01621459.1987.10478550.

Marron, J. S., and M. P. Wand. 1992. “Exact Mean Integrated Squared Error.” *Ann. Statist.* 20 (2): 712–36. doi:10.1214/aos/1176348653.

For defining \(\hat g(y;h)\) we consider the transformed kde seen in Section 2.5.1.↩

Since \(n^{-4/5}=o(n^{-2/3})\).↩

Not to confuse with

`bw.nrd0`

!↩Why so? Because \(X\) is unbiased for estimating \(\mathbb{E}[X]\).↩

Recall that another way of writing (2.28) is as the relative error \(\frac{\hat{h}-h_\mathrm{MISE}}{h_\mathrm{MISE}}\) being asymptotically distributed as a \(\mathcal{N}\left(0,\frac{\sigma^2}{n^{2\nu}}\right)\).↩