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First include the mean-gain calculation preceding the numbered clauses. At each electron-multiplying CCD stage one electron contributes an expected $1+p$ electrons to the next stage. If $N_j$ is the charge after stage $j$, conditional expectation gives $\mathbb{E}[N_{j+1}\mid N_j]=(1+p)N_j$. Iteration from $n$ input electrons yieldsFor small $p$, $g\simeq e^{rp}$ if corrections of order $r{p}^2$ are negligible.

In the high-gain approximation the output from one input electron has an exponential distribution of scale $g$. Independent input electrons produce independent cascades, and their charges add. By convolution of independent random variables, assuming the formula for $n$ inputs,Together with the single-electron base case this provesIt is a gamma distribution with shape $n$ and scale $g$, so it is normalized and has expected value $ng$ and variance $n{g}^2$. The formula is a continuous approximation to discrete output charge, not an exact integer-valued branching distribution; for $n=0$ there is a point mass at zero instead.

The deterministic equation has solution $n(t)=n(0)e^{(B-D)t}$ and is identical in form to the equation for the stochastic expected value. It therefore identifies only the net rateThe variance equation also contains the total event rate $s=B+D$. Once $r$ and $s$ are inferred,The deterministic trajectory cannot distinguish frequent births and deaths that nearly cancel from infrequent events with the same net growth, whereas fluctuations can.

For a uniform set family $\mathcal{A}\subseteq [n{]}^{(r)}$, the Local LYM inequality isTo prove it, count the incident pairs $(B,A)$ with $B\in \partial \mathcal{A}$, $A\in \mathcal{A}$, and $B\subset A$. Every $A$ contributes exactly $r$ pairs, whereas every $B$ belongs to at most $n-r+1$ members of $[n{]}^{(r)}$. Thereforewhich is the displayed inequality after using the binomial coefficient identity $r\left(\genfrac{}{}{0ex}{}{n}{r}\right)=(n-r+1)\left(\genfrac{}{}{0ex}{}{n}{r-1}\right)$.

The LYM inequality says that every antichain $\mathcal{A}\subseteq \mathcal{P}([n])$ satisfiesFor the proof from the local inequality, take the lowest occupied level below the middle and replace that level by its upper shadow. The result remains an antichain: any new containment involving a set in the upper shadow would already give a containment involving the member of $\mathcal{A}$ immediately below it. The dual form of the Local LYM inequality says that this replacement cannot decrease the Lubell mass. Repeating upward below the middle, and similarly replacing high levels by their lower shadows, eventually puts the whole family in one middle level. Its final Lubell mass is at most one, so the original mass is also at most one.

For the maximal chain in a Boolean lattice proof, choose a Uniformly random maximal chain in a Boolean lattice. An $h$-element set lies on it with probability $1/\left(\genfrac{}{}{0ex}{}{n}{h}\right)$. Because an antichain meets each chain at most once, the expected value of the number of its members on the chain is at most one. By linearity of expectation, that expected value is precisely the displayed sum.

The Sperner theorem follows because $\left(\genfrac{}{}{0ex}{}{n}{|A|}\right)\le \left(\genfrac{}{}{0ex}{}{n}{\lfloor n/2\rfloor }\right)$ for every $A$. If equality holds in the resulting cardinality bound, every member lies on a largest level. For even $n$ this is the unique middle level. For odd $n$, the two middle levels have equal size; the regular connected inclusion graph between them and equality in Local LYM inequality force a chosen portion of the lower level to be either empty or the whole level. Hence the maximum antichains are exactly the complete middle level, with either middle level allowed when $n$ is odd.

The Cameron-Martin space of a Gaussian measure $\mu =\mathcal{N}(0,\Sigma )$ isThe inverse is defined on the range of ${\Sigma }^{1/2}$. In this infinite-dimensional setting “positive definite” must mean $⟨\Sigma h,h⟩>0$ for every nonzero $h$, rather than a uniform lower bound: a trace-class operator cannot be uniformly positive on an infinite-dimensional Hilbert space.

By the spectral theorem for compact Hermitian operators, choose an orthonormal basis $(e_j)$ with $\Sigma e_j={\lambda }_j{e}_j$, ${\lambda }_j>0$, and ${\sum }_j{\lambda }_j<\infty$. (A strictly positive trace-class operator also forces the ambient Hilbert space to be separable.) In coordinates $h_j=⟨h,e_j⟩$,This is a Hilbert space with its indicated norm, even though its range need not be closed in the ambient norm.

The Cameron-Martin theorem for a Gaussian measure states that the translated law ${\mu }_h(B)=\mu (B-h)=\mathcal{N}(h,\Sigma )(B)$ is an equivalent probability measure to $\mu$ exactly when $h\in E$. For such $h$, its Radon-Nikodym derivative isThe series has mean-square convergence and converges almost surely under $\mu$, since its independent summands have total variance ${\sum }_j{h}_j^2/{\lambda }_j$. Its law is $\mathcal{N}(0,\Vert h{\Vert }_E^2)$, so the density has expected value one. It is sometimes formally written ${\ell }_h(x)=⟨h,x{⟩}_E$, but a typical infinite-dimensional Gaussian sample is not in $E$; the series interpretation is essential. If $h\notin E$, the two laws are mutually singular measures.

Choose the common dominating probability measure $\nu =(\mu +{\mu }^{\prime })/2$ and let $p=d\mu /d\nu$, $q=d{\mu }^{\prime }/d\nu$ be the Radon-Nikodym derivatives. The assumed second moments imply first-moment integrability by the Cauchy-Schwarz inequality. Since $Y$ is a separable Banach space, the measurable $f$ is a strongly measurable function, and both expected values exist as Bochner integrals.

The difference of these Bochner integrals satisfiesThe first inequality is the norm bound for a Bochner integral, and the second is the scalar Cauchy-Schwarz inequality. Since $(\sqrt{p}+\sqrt{q}{)}^2\le 2(p+q)$, the first factor is at mostWith the Hellinger distance convention of the original PDF, the second factor is $\sqrt{2}{d}_{\mathrm{Hell}}(\mu ,{\mu }^{\prime })$. Multiplication proves the Hellinger bound for differences of expectations:The proof uses only the two laws and their common dominating measure; no absolute continuity of measures assumption between $\mu$ and ${\mu }^{\prime }$ is needed.

Assume (ii). A martingale has constant expected value by the tower property of conditional expectation, so for every stopping time $\tau$ and every $n$,This is (iii).

Nash's theorem says that every finite game, including a two-player game with two pure choices each, has at least one Nash equilibrium in mixed strategies.

For player $1$, holding $p_2$ fixed, the expected value of the payoff isIt is an affine function of $p_1$. Therefore player $1$ chooses $p_1=1$ when $1-(U-1)p_2>0$, chooses $p_1=0$ when it is negative, and is indifferent when it vanishes. By symmetry, player $2$ has the analogous best response.

If $U<2$, then $1-(U-1)p>0$ for every $p\in [0,1]$, so becoming a Dark Lord with probability one is a strictly dominant strategy. The unique equilibrium isIf $U>2$, each player is indifferent when the other chooses $p=1/(U-1)$. This gives one interior mixed equilibrium, while the two asymmetric pure equilibria arise because a player facing a certain peasant chooses Lord and a player facing a certain Lord chooses peasant. The three equilibria areConsequently $\boxed{U_0=2}$. At the threshold itself there is a continuum of equilibria with $p_1=1$ or $p_2=1$, consistent with the question's strict inequalities.

For a fixed sample $z_{1:n}$, its Empirical Rademacher complexity isIts sample expected value is ${\mathcal{R}}_n(\mathcal{F})$.

Because every $f\in \mathcal{F}$ takes values in $\{0,1\}$, replacing one sample point changes $\hat{\mathcal{R}}$ by at most $1/n$. PutThe lower-tail form of the Bounded differences inequality givesoutside an event of probability at most $\delta /3$.

For the excess-loss supremum $G$ from part (b), replacing one observation changes $G$ by at most $2/n$. A second application of the same inequality givesoutside another event of probability at most $\delta /3$. On the intersection of these two events, symmetrization and the union bound yieldThis event has probability at least $1-2\delta /3\ge 1-\delta$, and $4s=2\sqrt{2\log (3/\delta )/n}$. Thus

The distribution is the geometric distribution on $\{1,2,\dots \}$ with success parameter $q=e^{-\lambda t}$. Its expected value is thereforeEquivalently, multiplying the master equation by $n$ and summing gives $m^{\prime }(t)=\lambda m(t)$ with $m(0)=1$.

Normalization of the probability density function givessoThe expected value isThe second moment isTherefore the variance is

For the median $m$,so

Expanding around the expected value $\mu$,because $\mathbb{E}[X-\mu ]=0$. Thereforewith equality exactly when $a=\mu$.

For the absolute loss,Leibniz differentiation gives, with $F(a)={\int }_{-\infty }^{a}f(x)dx$,Thus $H$ decreases while $F(a)<1/2$ and increases while $F(a)>1/2$. It is minimized at any median, characterized in the continuous case by

For points $x_1,\dots ,x_n$ and coefficients $c_1,\dots ,c_n$,The bound $|\hat{\varphi }|\le M$ ensures that this expected value is finite. Symmetry is immediate, so an expected outer product of a random feature map always defines a positive-semidefinite kernel.

Subtracting the expected value of $f$ does not alter either side, so suppose $\pi (f)=0$. Since reversibility makes $K$ a self-adjoint operator and stationarity makes it a contraction,The Discrete-time Poincaré inequality for a Markov kernel is therefore equivalent toApplying this inequality successively to $f,Kf,\dots ,K^{t-1}f$ yieldsConversely, the asserted variance contraction with $t=1$ rearranges to the Poincaré inequality. Hence the two statements are equivalent.

Write $p=(1+\mu {)}^{-1}$. Since $X$ and the geometric random variable $Z$ have the same expected value $\mu$,This is also the entropy deficit relative to the maximum entropy distribution on the nonnegative integers.

The random variables $X$ and $-Z$ are independent, so part a gives $H(X-Z)\ge H(-Z)=H(Z)$. Consequentlywhich is the required bound in terms of the Entropic Ruzsa distance.

To first order in the weak fluctuation, $n^2-1=2\mu W+O({\mu }^2)$, soForintroduce the signs $s=(1,-1,1,-1)$. The product rule gives

Assume that $W$ is a zero-mean stationary Gaussian random field, that its longitudinal correlation length is short compared with the envelope's evolution scale, and that the propagation distance is long compared with that correlation length. The forward Markov approximation then neglects diffraction during one correlation length. Applying the Furutsu–Novikov formula closes the last average at second order in $\mu$. Define the integrated longitudinal autocorrelation function of a random fieldThenwhereWriting $C_{ij}=C(z_i-z_j)$ and using the evenness of the covariance gives the equivalent expressionThe derivation also assumes paraxial propagation, weak scattering, sufficient regularity to interchange differentiation and expectation, and statistical homogeneity in both coordinates. Without the short-correlation approximation, the Gaussian identity produces a nonlocal longitudinal memory integral rather than this local closed equation.

Let $X$ count copies of the complete graph $K_4$ in $G$. Its expected value is

Write $X$ as a sum of indicator random variable. Indicators for two distinct copies are independent when the copies share at most one vertex, since their edge sets are then disjoint. Pairs sharing two vertices have eleven edges in their union, while pairs sharing three have nine. It follows thatThus $\mathrm{var}(X)/{\mu }^2\to 0$. By the second moment method, $X/\mu \to 1$ in probability, so $\mathbb{P}(X\ge 100)\to 1$. This is the sparse clique-count concentration estimate.

Condition on the information available before time $n$. The current wealth is then fixed, while ${\xi }_n\sim N(b,\Sigma )$ is independent of that information. PutFor any portfolio $\theta$, use the inner product induced by the positive-definite matrix $\Sigma$ to writeThe linear image of a multivariate normal vector shows that the two portfolio returns are jointly normal, andThus independence of uncorrelated jointly normal variables makes the residual return ${\eta }^T{\xi }_n$ independent of the aligned return; it also has expected value zero.

The continuation value $V(n,\cdot )$ is concave by part (c). Conditional Jensen inequality therefore shows that adding the independent centered residual cannot improve the objective:where $m=(1+r)X_{n-1}$. Since the optimal portfolio is unique, its residual must be zero.

It remains to determine the sign. If $\lambda <0$, the portfolios $\lambda {\Sigma }^{-1}b$ and $-\lambda {\Sigma }^{-1}b$ have the same return variance, namely ${\lambda }^{2}B$, while their means are $\lambda B$ and $-\lambda B$. The latter return has the distribution of the former plus the positive constant $-2\lambda B$. Since $V(n,\cdot )$ is increasing, replacing $\lambda$ by $-\lambda$ cannot reduce the objective, contradicting uniqueness. Hence $\lambda \ge 0$.

Applying this conditional argument at every time gives nonnegative, past-measurable random variables ${\lambda }_n$ such thatThis is the Gaussian one-fund theorem.

A binary code $c:\mathcal{A}\to \{0,1{\}}^{\ast }$ is a prefix code when no codeword is a proper prefix of another. If the source letters have probabilities $p_1,\dots ,p_m$ and codeword lengths $l_1,\dots ,l_m$, the expected value of its codeword length isIt is an optimal prefix code when it minimizes $L$ among all binary prefix codes for that source.

Suppose $p_i>p_j$ but $l_i>l_j$. Exchanging the codewords assigned to ${\mu }_i$ and ${\mu }_j$ leaves the set of words, and hence prefix-freeness, unchanged. The change in expected length iscontradicting optimality. Thuswhich is the probability monotonicity of optimal prefix-code lengths.

Represent the code by its binary prefix tree and choose a codeword $wb$ of maximal length, where $b\in \{0,1\}$. Its sibling is $w(1-b)$. If that sibling were not a codeword, it could not have a codeword below it either, since such a word would be longer than $wb$. Replacing $wb$ by $w$ would then preserve prefix-freeness and strictly reduce $L$. This is impossible for an optimal code, so $w(1-b)$ is also a maximal-length codeword. The pair differs only in its last digit, proving the deepest sibling property of an optimal prefix code.

With population $n$, the transition diagram has the two outgoing arrowsThus this is a batch-birth linear-death process. If $P_n(t)=\mathbb{P}(N_t=n)$ and $P_j=0$ for $j<0$, probability enters state $n$ from $n-3$ by a birth and from $n+1$ by a death. Hence the master equation is

The Markov jump-process generator isFor $f(n)=n$, this gives $Ln=3\lambda -\beta n$. Therefore the expected value $m(t)=\mathbb{E}[N_t]$ satisfiesso

For $f(n)=n^2$,Writing $s=\mathbb{E}[N_t^2]$ and using the generator identity givesThe variance $v=s-m^2$ consequently obeysIf $v(0)=v_0<\infty$, substitution of the formula for $m$ yieldsAll transient terms vanish, and the moments of a batch-birth linear-death process therefore give

The mean extension in an ensemble of $A$ molecules iswith constraints ${\sum }_i{a}_i=A$ and ${\sum }_i{a}_i{l}_i=AL$. For occupancies $\{a_i\}$, the number of assignments of the $A$ distinguishable ensemble members to the states is the multinomial coefficientThus maximizing the probability at fixed $A$ and $L$ is equivalent to maximizing the stated Lagrange multiplier expression.

Because $A\gg 1$, the Stirling formula gives $\log (a_i!)\simeq a_i\log a_i-a_i$. Differentiating with respect to each $a_i$ givesNormalization by ${\sum }_i{a}_i=A$ therefore yields the probability mass functionHere $Z$ is the fixed-tension partition function that normalizes the probabilities.