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The all-ones matrix $J=\mathbf{e}{\mathbf{e}}^T$ has every entry equal to one, where $\mathbf{e}=(1,\dots ,1{)}^T$. On ${\mathbb{R}}^n$, the matrix $J/n$ is the orthogonal projection matrix onto $\mathrm{span}\{\mathbf{e}\}$.

If a symmetric matrix $H$ has largest eigenvalue ${\lambda }_1$ separated from ${\lambda }_2$ by $\gamma >0$, and $P=v{v}^{\top }$ is its leading rank-one orthogonal projection matrix, then every unit vector $u$, with $Q=u{u}^{\top }$, satisfies the displayed inequality. Expanding $u$ in an orthonormal eigenbasis gives ${\lambda }_1-u^{\top }Hu\ge \gamma (1-|u^{\top }v{|}^2)$, while $\Vert P-Q{\Vert }_F^2=2(1-|u^{\top }v{|}^2)$. The analogous statement for a smallest eigenvalue follows by replacing $H$ by $-H$.

For an integer $0\le k\le n$, the fantope is the convex hull of rank-$k$ orthogonal projection matrices in ${\mathbb{R}}^n$. Equivalently, it consists of real symmetric matrices with eigenvalues in $[0,1]$ and matrix trace $k$. To prove the equivalence, apply the spectral theorem for real symmetric matrices and express the eigenvalue vector as a convex combination of the zero-one extreme points of the capped simplex. The sum of the largest eigenvalues is its support function.

The Frobenius norm is the finite-matrix Hilbert-Schmidt norm. It equals $\sqrt{\mathrm{tr}(A^{\ast }A)}$, is unchanged by unitary multiplication on either side, and is the square root of the sum of squared singular values. For unit vectors $u,v$, $\Vert u{u}^{\ast }-v{v}^{\ast }{\Vert }_F^2=2(1-|u^{\ast }v{|}^2)$, a useful sign-invariant measure of distance between rank-one orthogonal projection matrices.

For a real symmetric matrix $X$ and $1\le k\le n$,Indeed $Y=Q{Q}^T$ is a rank-$k$ orthogonal projection matrix in the fantope, and projecting onto the eigenvectors of the largest eigenvalues attains the support function maximum. Equivalently, the largest matrix trace of a compression to a $k$-dimensional vector subspace is the sum of the largest eigenvalues.

The mean-vector prediction risk measures squared Euclidean norm error in estimating the deterministic response mean. For independent future noise with covariance matrix ${\sigma }^2{I}_n$, prediction of a new noisy response adds $n{\sigma }^2$ to this risk. A Gaussian orthogonal projection matrix $P$ of rank $k$ has risk $\Vert (I-P)\mu {\Vert }^2+k{\sigma }^2$.

Full column matrix rank makes $X^{T}X$ a positive-definite matrix, and its matrix inverse is symmetric. Thus the hat matrix satisfiesAlso $PX=X$ and the image of $P$ is contained in the column space of $X$. These identities show that $P$ is the orthogonal projection matrix onto the column space of $X$.

Represent the subset model by a set $M$ of $k$ column indices and let $X_M$ contain those columns. Its ordinary least squares estimator sets the omitted coefficients to zero and hasFor $k=0$, use $P_M=0$. Full column rank of $X$ implies full column rank for each $X_M$. The matrix $P_M$ is an orthogonal projection matrix of rank $k$.

Put $\mu =X\theta$ and $b_M=(I-P_M)\mu$. Projection orthogonality gives the bias-variance decomposition of mean squared errorwhile the expected residual sum of squares isThese equations distinguish the mean-vector prediction risk in this problem from the additional noise in predicting a new response vector.

First take $p<n$. Let $P_X$ project onto the full column space of $X$. Since the full model contains $\mu$, the residual estimate of Gaussian noise varianceis unbiased: its numerator has expectation $(n-p){\sigma }^2$. Consequently an unbiased Gaussian projection risk estimate isIndeed its expectation is $\Vert b_M{\Vert }^2+(n-k){\sigma }^2+(2k-n){\sigma }^2=R(M)$ for every $\theta$, even when the subset model omits true effects. Independence of the two terms is not needed for this expectation calculation. The full-model variance estimate is important: the subset residual variance generally includes omitted-variable bias.

For a finite candidate collection, choose a model minimizing $\hat{R}(M)$, breaking ties in favour of a smaller model. The term $-n{\hat{\sigma }}^2$ is common to all models, so this amounts to minimizingthe Mallows Cp form. The fit improvement competes with an increasing dimension penalty. Unbiasedness holds for each fixed model; the estimate at the data-selected model need not remain unbiased, because selection favours downward fluctuations. The method is a heuristic model selection rule, rather than an assertion that it finds the true model with certainty.

The printed $p\le n$ includes a genuine qualification. If $p=n$, there are no full-model residual degrees of freedom. Except in the balanced case $2k=n$, no integrable data-only unbiased estimator can satisfy the requested identity for all means and unknown variances. Here is a proof of the unknown-variance risk estimation in a saturated Gaussian model obstruction. Since $X$ is invertible, $\mu$ ranges over all ${\mathbb{R}}^n$. Suppose $T(Y)$ had the required expectation at every $\mu$ and variance $s>0$. For $v>0$, let $U\sim N_n(\mu ,vI)$ and let $Y\mid U\sim N_n(U,sI)$. Marginally $Y\sim N_n(\mu ,(s+v)I)$. Applying the assumed identity conditionally giveswhereas applying it to the marginal distribution gives $\Vert b_M{\Vert }^2+k(s+v)$. Their difference is $(n-2k)v$, which must vanish. Integrability under the marginal Gaussian justifies conditioning. If $2k=n$, the obstruction disappears and the residual sum of squares alone has expectation $\Vert b_M{\Vert }^2+k{\sigma }^2$. ThusThe general construction therefore requires $p<n$, or an independently available unbiased variance estimate.

Write the one-column design as $x\in {\mathbb{R}}^n\setminus \{0\}$ and $S=x^{T}x$. Maximizing the normal linear model likelihood first gives $\hat{\beta }=x^{T}Y/S$. Its minimized residual sum of squares is $R=\Vert Y-x\hat{\beta }{\Vert }^2$. For variance $v>0$, the remaining log likelihood is $-\frac{n}{2}\log v-R/(2v)$ plus a constant, whose maximum is atHere $R>0$ almost surely because $n>1$ and the true variance is positive. In the exceptional data case $R=0$, the likelihood is unbounded as $v\downarrow 0$, so there is no positive-variance maximum.

Let $P=x{x}^T/S$, the orthogonal projection matrix onto the design column. Then $\hat{\beta }-\beta =x^T\epsilon /S$ and $Y-x\hat{\beta }=(I-P)\epsilon$. These are orthogonal linear projections of an isotropic multivariate normal distribution, hence independent. An orthonormal coordinate system adapted to $x$ makes the residual squared length a sum of $n-1$ independent squared standard normals. Thus the joint distribution of least-squares and variance estimators is the product lawIn particular the maximum likelihood variance estimator is biased: $\mathbb{E}{\hat{\sigma }}^2=(n-1){\sigma }^2/n$; the unbiased residual estimator uses denominator $n-1$.

Let $P=u{u}^T/\Vert u{\Vert }^2$. This is an orthogonal projection matrix, so $P^T=P$ and $P^2=P$. Therefore the Householder transformationsatisfies $H_u^T{H}_u=(I-2P{)}^2=I$. If $\Vert a\Vert =\Vert b\Vert$ and $a\ne b$, thenand hence $H_{a-b}a=a-(a-b)=b$.

Apply a Householder transformation to the first column of $A$ to map it to a multiple of $e_1$, then apply transformations supported on the trailing coordinates to clear each later column below its diagonal. Their product is orthogonal and produces $R$; reversing the product gives $A=QR$.

For the displayed matrix, one resulting factorization isDirect multiplication gives the stated $A$, and $Q^{T}Q=I$.

Replace coordinate inequalities by the Loewner order and coordinate sums by matrix trace. The maximizing semidefinite program isIts feasible set is the fantope. In an orthonormal eigenbasis of $X$, its objective is ${\sum }_i{\lambda }_i(X)Y_{ii}$, and the diagonal entries of $Y$ belong to the capped simplex. Choosing $Y$ as the orthogonal projection matrix onto the largest $k$ eigenvectors attains the sum of the largest eigenvalues, as in the Ky Fan maximum principle.

The minimizing semidefinite program isFor example, its Lagrangian arises by assigning $S⪰0$ to $I-Y⪰0$, keeping $Y⪰0$ in the domain, and using free $t$ for $\mathrm{tr}Y=k$. Finiteness of the supremum over $Y$ requires $S+tI-X⪰0$.

Equality follows directly by choosing $S$ to have eigenvalues $({\lambda }_i(X)-t{)}_{+}$ in the same orthonormal eigenbasis, with $t$ between the $k$th and $(k+1)$st eigenvalues, or below the smallest one for $k=n$. This is the threshold semidefinite program for the largest eigenvalues, and it also verifies the endpoint case $k=n$.

Because $\mathrm{col}Z\subseteq \mathrm{col}X$, the two orthogonal projection matrices satisfyConsequently $P-P_0$ is itself an orthogonal projection, onto the orthogonal complement of $\mathrm{col}Z$ within $\mathrm{col}X$. In particular it is a positive semidefinite matrix, so for every $u\in {\mathbb{R}}^n$,

For an orthogonal projection matrix $P=P^{†}=P^2$,Thus $P$ is a positive semidefinite matrix.