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The expected value and variance under quota share reinsurance follow by scaling the exponential distribution:The retained stop loss moments for an exponential aggregate follow from the payout $Y=\min (S,M)$, its survival function equals $e^{-x/{\mu }_S}$ for $0\le x<M$ and zero for $x\ge M$. The tail integral formula for moments gives, with $m=M/{\mu }_S$ and $r=e^{-m}$,Consequently the retained moments areMatching the two expected values forces $\alpha =1-r$, which lies strictly between zero and one. The difference of the variances simplifies toIndeed $h(m)=m-1+e^{-m}$ has $h(0)=0$ and $h^{\prime }(m)=1-e^{-m}\ge 0$ for $m\ge 0$. Because $M>0$, the difference is actually positive. At equal retained expected value, aggregate stop loss reinsurance reduces the variance more than quota share reinsurance.

Write $Y=g(S)$ and $Z=\min (S,M)$. Their common expected value is $c$. Expanding about the retention gives the requested variance identityThe stop loss variance minimization principle follows from a pointwise comparison. For $0\le x\le M$, the constraint $0\le g(x)\le x$ impliesFor $x>M$, the squared distance of $\min (x,M)=M$ from $M$ is zero, so the same squared-distance comparison is immediate. ThereforeSubtracting the same $(M-c{)}^2$ proves optimality of the retained stop loss payout:Since $Z$ is bounded, its variance is finite; if $\mathbb{E}[Y^2]=\infty$, the inequality remains valid with infinite variance on the left. When it is finite, equality requires $g(S)=\min (S,M)$ almost surely, because the pointwise squared-distance inequality is strict whenever the two payouts differ.

Use $c_{\mathrm{prem}}$ for the premium income rate, reserving $c$ later for the smaller exponential decay rate. In the classical risk model, the surplus isHere $\rho$ is the relative safety loading. The aggregate claims form a Compound Poisson process. Define $L_t={\sum }_{j=1}^{N_t}{X}_j-c_{\mathrm{prem}}t$, so ruin occurs when $L_t>u$. By independent increments and the exponential formula for a marked Poisson sum,Thus $Z_t=e^{R{L}_t}$ is a nonnegative continuous-time martingale with $Z_0=1$, because the adjustment coefficient makes the exponent vanish.

Let $\tau =\inf \{t:U_t<0\}$. Apply the optional stopping theorem at the bounded stopping time $\tau \wedge T$. On $\{\tau \le T\}$, $L_{\tau }>u$, soLetting $T$ increase proves the Lundberg inequality and the unscaled limit:

For the precise asymptotic, putThe given exponential integral identity makes $k$ a probability density. Multiplying the given defective renewal equation by $e^{Ru}$ turns it into the ordinary renewal equationFor clarity, the version of the key renewal theorem used here is: if the interarrival law is nonarithmetic, has mean $m_k\in (0,\infty )$, and $g$ is directly Riemann integrable, the locally bounded solution of this renewal equation satisfies $z(u)\to m_k^{-1}{\int }_0^{\infty }g(v)dv$. The infinite-mean version gives zero for nonnegative directly Riemann integrable $g$.

All the hypotheses can be checked here. The density $k$ gives a nonarithmetic distribution. The Tonelli theorem givesFurthermoreThus $g$ is continuous and integrable, and ${\int }_0^{\infty }|g^{\prime }(u)|du\le R\int g+1<\infty$. On a mesh of width $\delta$, the difference between its upper and lower sums is at most $\delta \int |g^{\prime }|$; its upper sum is at most $\int g+\delta \int |g^{\prime }|$. This proves direct Riemann integrability rather than assuming it. Also $0\le z(u)\le 1$ by the Lundberg inequality, so the solution is locally bounded. Its renewal representation is $z=g\ast U_k$, where $U_k={\sum }_{j\ge 0}{K}^{\ast j}$ and $K(dx)=k(x)dx$; the residual after iteration tends to zero on compact intervals because sums of positive interarrivals tend to infinity.

Writing $J={\int }_0^{\infty }x{e}^{Rx}{f}_I(x)dx$, the tilted interarrival expected value is $m_k=hJ$. The key renewal theorem gives the Cramér–Lundberg ruin asymptoticIf $J=\infty$, the same formula is interpreted as $A=0$. A positive finite asymptotic constant requires $J<\infty$; this extra integrability is not explicitly stated in the paper.

For the final two-exponential case, evaluate the defective renewal equation at zero:One can identify the adjustment coefficient without silently assuming $A>0$. For $0<r<c$, setIt is finite and positive. Integrating the nonnegative terms of the defective renewal equation, using the Tonelli theorem, first shows that $F_I(r)={\int }_0^{\infty }{e}^{rx}{f}_I(x)dx$ is finite and then givesAs $r\uparrow c$, $P(r)\to \infty$ because $a>0$. By monotone convergence theorem, $h{F}_I(c)=1$. The integrated tail distribution in the classical risk model has density $f_I(x)=\mathbb{P}(X_1>x)/\mu$, so $F_I(r)=[M(r)-1]/(\mu r)$ by the tail integral formula for moments. Hence $c$ solves the adjustment equation, and its stipulated uniqueness implies $R=c$. Finally the displayed form of $\psi$ gives the remaining constantsIn particular the decay exponent $d$ and the coefficient $b$ do not affect $R$ or $A$. The $c$ in these final answers is the printed decay rate, not $c_{\mathrm{prem}}$.

In the Bühlmann model, a latent risk parameter $\Theta$ is drawn from a population distribution. Conditional on $\Theta$, the yearly observations are independent and identically distributed random variables, with conditional expectation $m(\Theta )$ and conditional variance $v(\Theta )$. Define the structural parametersHere $v$ is the expected process variance, while $a$ is the variance of hypothetical means. The Bühlmann credibility premium is the best affine estimate of $m(\Theta )$ from the observed claims, under mean squared error. Predicting the next claim gives the same affine estimate: the extra conditional observation noise contributes the constant $v$ to the prediction error.

The law of total variance and conditional independence giveAn affine estimate can be written as $\hat{m}=m+{\sum }_{j=1}^n{b}_j(X_j-m)$: for any chosen $b_j$, optimizing the constant makes its expected value equal to $m$. The normal equations for the linear least-squares projection areFor $v>0$ they force all $b_i$ to agree, with $b_i=a/(v+na)$. Thus the credibility factor and premium areThe credibility factor increases with the observation count and between-risk variance, and decreases with within-risk variance. If $a=0$, the risk mean is known and $Z=0$; if $v=0$ and $a>0$, one observation reveals it and $Z=1$. If both vanish, the premium is the fixed value $m$ and the factor is immaterial.

In the specified model, the conditional law is a gamma distribution with shape $\alpha$ and scale $\theta$. ThereforeThe prior is an inverse-gamma distribution with shape $k$ and scale $\lambda$. To obtain its moments directly, substitute $y=\lambda /\theta$ in the defining integral, obtainingThe Gamma function recurrence yieldsThe assumption $k>2$ makes both structural variances finite. HenceIt follows that the model-specific credibility estimate is

For the Bayes estimator under squared error loss, the quantity to estimate is $\alpha \Theta$, so the optimum is its posterior mean. The likelihood function, viewed as a function of $\theta$, is proportional toMultiplication by the prior shows gamma scale inverse-gamma conjugacy:Although the printed hint only mentions integer shapes, the same substitution and Gamma integral normalize this posterior for every positive real shape, so no integrality of $\alpha$ is needed. Its posterior mean gives the Bayesian estimate and comparisonThis exact Bühlmann credibility for gamma claims holds for every observed sample, not merely on average. Here the posterior mean is affine in the sample mean, so the best affine Bühlmann credibility premium is also the unrestricted Bayes estimator under squared error loss.

The expected values of early and late new diagnosis counts at step $k$ areEach summand in $n_k$ corresponds to an infection that avoids early diagnosis, enters the non-early state one step later, remains there for the intervening steps and is then diagnosed. Define $h_j=0$ for $j\le 0$, and take an empty sum as zero. The early share of diagnosis intensity isIt is defined only when $e_k+n_k>0$; in particular there are no diagnoses at step 1 under the initial condition. With infection counts following a Poisson distribution and independent marking, early and late diagnosis counts in a fixed interval are independent Poisson random variables. Conditional on a positive total, the early count has a binomial distribution given the total with probability $p_{Ek}$, so this ratio also equals the expectation of the observed early fraction conditional on a nonzero total. Without that conditioning a sample fraction at zero diagnoses is undefined.

A noncausal stationary autoregression exists. On the two-sided time axis defineThis series converges in the sense of mean-square convergence, and direct subtraction gives $X_t-2X_{t-1}={\epsilon }_t$. The expected value and autocovariance areConsequently this is a weakly stationary process and, by Gaussianity, a strictly stationary process. It is an example of noncausal stationary autoregression, since $X_t$ uses future noise. In particular, $X_{t-1}$ is correlated with ${\epsilon }_t$, so the usual causal variance recursion is inapplicable. If an additional assumption required the driving noise to be independent of past observations, this solution would be excluded; that assumption is not stated here.

A stationary causal time series exists. The infinite moving-average representationconverges in the sense of mean-square convergence because the driving variables are independent random variables and ${\sum }_j{4}^{-j}<\infty$. Shifting the series gives $X_t=\frac{1}{2}{X}_{t-1}+{\epsilon }_t$. Its expected value is zero and its autocovariance isThus it is a weakly stationary process; since the driving variables have a normal distribution, it is also a Gaussian process and a strictly stationary process. Throughout the stationarity discussion, take ${\sigma }^2>0$; degenerate zero noise permits trivial constant solutions.

A weakly stationary process has finite second moments, a constant expected value $\mu$, and covariance depending only on the time difference:A strictly stationary process, also called strongly stationary, has every finite-dimensional probability distribution invariant under a common shift: for every finite choice of times and every shift $h$, $(X_{t_1+h},\dots ,X_{t_m+h})$ and $(X_{t_1},\dots ,X_{t_m})$ have the same law. strict stationarity with finite second moments implies weak stationarity. weak stationarity alone does not determine the full joint law, and strict stationarity alone does not guarantee finite moments.

First justify the dependence structure rather than assume that adjacent squares behave like an ordinary lag-one ARCH model. Write $Z_t=X_t^2$. Iterating its nonnegative recurrence givesThe product coefficient has expected value ${\alpha }_2^m$, so it tends to zero in probability. The stationary $Z_{t-2m}$ are a tight family; consequently the remainder tends to zero in probability, without needing independence between that remainder's factors. The increasing partial sums therefore give the representationThis proves the parity decomposition of a lag-two ARCH process: even and odd observations are functions of disjoint noise families. It also shows that the positive scale ${\sigma }_t$ is a function of past noise, independent of ${\epsilon }_t$. The conditional expectation of $X_t$ given past noise is zero, and the second-moment recursion is $m_2={\alpha }_0+{\alpha }_2{m}_2$. independence of the parity families gives the adjacent-square covariance. ThusTo establish finiteness of the fourth moment before using its recursion, note that the $L^2$ norm of the $k$th term in the positive series for $Z_t$ is ${\alpha }_0\sqrt{3}({\alpha }_2\sqrt{3}{)}^k$. The triangle inequality in L2 space makes the series square-integrable when $3{\alpha }_2^2<1$. independence of the current noise and past scale then yieldsand henceIf $3{\alpha }_2^2\ge 1$, a finite $m_4$ would make $(1-3{\alpha }_2^2)m_4=3{\alpha }_0^2+6{\alpha }_0{\alpha }_2{m}_2>0$, which is impossible. Its fourth moment is then infinite. The adjacent-square product remains integrable by parity independence, even in that case.

A convenient general-state-space ergodic theorem is the following. For a positive Harris recurrent Markov chain with invariant probability measure $\pi$, and a measurable function $h$ with $\pi |h|<\infty$,Here $\pi h=\int hd\pi$ is the stationary expected value. Harris recurrence means that every set of positive irreducibility measure is visited almost surely from every state; positive recurrence supplies an invariant probability rather than only an infinite invariant measure. The ergodic theorem for a positive Harris recurrent Markov chain holds from any starting state under these Harris hypotheses. aperiodicity is not necessary just for averages.

A useful sufficient form of the central limit theorem for a geometrically ergodic Markov chain adds aperiodicity, geometric ergodicity, and $\pi (|h{|}^{2+\delta })<\infty$ for some $\delta >0$. It givesThese are sufficient hypotheses, not a claim that irreducibility alone ensures a central limit theorem. The Markov chain Monte Carlo asymptotic variance uses stationary covariances ${\gamma }_h(k)={\mathrm{Cov}}_{\pi }(h(X_0),h(X_k))$, with ${\gamma }_h(0)={\mathrm{Var}}_{\pi }h$. The series is absolutely convergent under the stated sufficient assumptions. If ${\gamma }_h(0)>0$, write ${\rho }_h(k)={\gamma }_h(k)/{\gamma }_h(0)$ and $v_h={\gamma }_h(0){\tau }_{\mathrm{int}}$, where the integrated autocorrelation time is ${\tau }_{\mathrm{int}}=1+2{\sum }_{k\ge 1}{\rho }_h(k)$. When $v_h>0$, the effective sample size of a Markov chain is approximately $n/{\tau }_{\mathrm{int}}$. Negative correlations can reduce the asymptotic variance; a zero asymptotic variance gives a degenerate normal limit. For constant $h$, variance is zero and the autocorrelation normalization is undefined.

Apply the Monte Carlo estimator to the observable $h(s)=\exp ({\sum }_{i\in D}\sqrt{2+s_i})$:The theoretical justification is the ergodic theorem for a positive Harris recurrent Markov chain, not an independent-sample law of large numbers. The Markov chain has finite state space, is irreducible, and has the strictly positive posterior distribution as invariant law, so it is positive recurrent and Harris recurrent with counting measure. The observable is bounded: $e^{K^2}\le h(s)\le e^{\sqrt{3}{K}^2}$. Thus the estimator converges almost surely to the posterior expected value from any initial configuration.

Each state also has positive self-transition probability, so the chain is an aperiodic Markov chain. Finiteness then gives geometric convergence and the relevant central limit theorem; serial dependence determines the Markov chain Monte Carlo asymptotic variance if error bars are required. A fixed finite burn-in can be discarded without changing consistency, but independence of the retained states should not be assumed.

Think of a vertex of this hypergraph as an edge of the complete graph on $[N]$. A hyperedge is the edge set of one $t$-vertex clique. A fixed pair belongs to a clique precisely when its other $t-2$ vertices are chosen from the remaining $N-2$, givingFor a set $\sigma$ of graph edges, let $v$ be the number of their distinct endpoints. If $v>t$, no $t$-clique contains them and the codegree is zero. Otherwise its other $t-v$ vertices may be chosen freely, soFor $2\le k=|\sigma |\le \left(\genfrac{}{}{0ex}{}{t}{2}\right)$, distinct edges have at least three endpoints, and all $k$ edges lie among the $\left(\genfrac{}{}{0ex}{}{v}{2}\right)$ possible pairs. Thus $3\le v\le t$ and $k\le \left(\genfrac{}{}{0ex}{}{v}{2}\right)$. In particular,This also identifies why the relevant exponent is $2/(t+1)$.

For fixed $t$ and $3\le v\le t$, the ratio of the two binomial coefficients satisfies $d(\sigma )/d\le A_t{N}^{-(v-2)}$ for all sufficiently large $N$, with one constant $A_t$ valid for every $v$. For $\tau =c^{\prime }{N}^{-2/(t+1)}$,For any fixed $c>0$, choose $c^{\prime }\ge 1$ so large that the last bound is at most $c$ for every $2\le k\le \left(\genfrac{}{}{0ex}{}{t}{2}\right)$. Zero codegrees satisfy the bound automatically. The asserted small-codegree condition therefore holds uniformly. Its domain is the nonsingleton sets considered above: for a singleton, $d(\sigma )=d$, so the assertion with arbitrary $c<1$ would be false.

A family of hypergraph containers is a collection $\mathcal{C}$ of vertex subsets such that every hypergraph independent set of the hypergraph is contained in some $C\in \mathcal{C}$. Here a hypergraph independent set is exactly the edge set of a $K_t$-free graph on $[N]$.

We use the following edge-sparse form of the hypergraph container theorem. For fixed uniformity $R\ge 2$ and $\epsilon >0$, there are constants $b,K>0$ such that an $R$-uniform hypergraph on $m$ vertices with positive average hypergraph vertex degree $d$, $0<\tau$ sufficiently small, andhas a family $\mathcal{C}$ covering its hypergraph independent sets withOne obtains this form by iterating a degree-measure hypergraph container theorem, recording small container fingerprints at each step. To ensure the requested strict edge inequality, apply it with $\epsilon /2$.

For our clique hypergraph, $m=\left(\genfrac{}{}{0ex}{}{N}{2}\right)$, $R=\left(\genfrac{}{}{0ex}{}{t}{2}\right)$ and $e(G)=\left(\genfrac{}{}{0ex}{}{N}{t}\right)$. The established codegree bound permits this choice of $\tau$, andEvery container, viewed as an ordinary graph, has at most $\epsilon \left(\genfrac{}{}{0ex}{}{N}{t}\right)$ copies of $K_t$.

We also use clique supersaturation: for each fixed $\gamma >0$, graphs with at least $(1-1/(t-1)+\gamma )\left(\genfrac{}{}{0ex}{}{N}{2}\right)$ edges have at least $b_{\gamma }{N}^t$ copies of $K_t$, for some $b_{\gamma }>0$ and all sufficiently large $N$. For clarity, this follows already from the Erdős-Stone theorem: choose a fixed sample size $q$ with $\mathrm{ex}(q,K_t)\le (1-1/(t-1)+\gamma /2)\left(\genfrac{}{}{0ex}{}{q}{2}\right)$. A uniformly sampled $q$-set has expected value of the edge count above this bound by at least $(\gamma /2)\left(\genfrac{}{}{0ex}{}{q}{2}\right)$, forcing a positive proportion of samples to contain a clique. Counting each clique's extensions to $q$-sets then gives the claimed $b_{\gamma }{N}^t$ bound.

Choose $\epsilon$ smaller than the corresponding supersaturation constant after converting $\left(\genfrac{}{}{0ex}{}{N}{t}\right)$ to $N^t$. Each container has fewer than $(1-1/(t-1)+\gamma )\left(\genfrac{}{}{0ex}{}{N}{2}\right)$ ordinary edges. Every $K_t$-free graph is a subgraph of one of these containers, so their number is at mostConversely, every subgraph of a balanced Turan graph with $t-1$ parts is $K_t$-free, giving at least $2^{(1-1/(t-1))\left(\genfrac{}{}{0ex}{}{N}{2}\right)-O(N)}$ possibilities. Letting $\gamma$ be arbitrarily small proves the enumeration formula:

Write $f(x)=\mathrm{TS}\{x_1,\dots ,x_n\}$. Besides Talagrand's convex distance inequality, we use the precise squared edge bound for tours in the unit square stated above and the triangle inequality, which permits shortcutting a closed walk without increasing its length. We first prove the geometric comparison that connects these results.

For each fixed $x$, choose a cyclic tour $C$ with edge lengths ${\ell }_i$ satisfying ${\sum }_i{\ell }_i^2\le 4$. At vertex $x_i$, let $a_i(x)$ be the sum of the lengths of its two incident edges in $C$. The Cauchy-Schwarz inequality givesWe claim the weighted mismatch bound for Euclidean tours:To prove it, keep the vertices whose coordinates agree. In the cyclic ordering $C$, the deleted vertices split into consecutive runs between kept vertices. For one run, let $a,b$ denote its two boundary edge lengths and $S$ the sum of its internal edge lengths. Starting at the left kept endpoint, visit the run and return along the same path, at cost $2(a+S)$; starting at the right endpoint costs $2(b+S)$. Choose the cheaper detour. Its cost is at mostwhich is exactly the sum of $a_i(x)$ over that run. Attach these detours to a shortest tour through $y$, which already visits every kept point. The resulting closed walk visits all points of $x$; shortcut it using the triangle inequality. Summing the detour costs proves the claimed weighted mismatch bound for Euclidean tours. If there is just one kept vertex, the same argument uses that vertex as both endpoints of the unique run. If there are no kept vertices, $f(x)\le {\sum }_i{\ell }_i\le {\sum }_i{a}_i(x)$ proves the comparison directly. Repeated point locations cause no difficulty, since their connecting edges may have length zero.

Let $M$ be any median of $f(X)$ and set $A=\{y:f(y)\le M\}$. For $f(x)\ge M+t$, the weighted mismatch bound for Euclidean tours impliesThe weights have squared sum at most one, so $d_T(x,A)\ge t/4$. Since $\mathbb{P}(A)\ge 1/2$, Talagrand's convex distance inequality yieldsFor the lower tail, put $B=\{x:f(x)\le M-t\}$. If $\mathbb{P}(B)=0$ there is nothing to prove. For every $y$ with $f(y)\ge M$, apply the same weighted mismatch bound for Euclidean tours with its weights $a_i(y)$ and all $x\in B$. It gives $d_T(y,B)\ge t/4$. The set of such $y$ has probability at least $1/2$, so Talagrand's convex distance inequality instead givesCombining both tails,The bound has a constant fluctuation scale, independent of $n$. Only independence of the point coordinates and their support in the square are needed here; their uniform distribution was needed for the earlier lower bound on the expected value.

Interpret $Y_n$ as the length of a shortest Euclidean travelling salesman tour, and take the points to be independent random variables with the uniform distribution on the square. We assume $n\ge 2$; a closed tour on one point has length zero, so the printed lower bound would be false for $n=1$. For two points the closed tour traverses the joining segment twice.

For the deterministic upper bound, use the following precise geometric result, the squared edge bound for tours in the unit square: every finite set of points in $[0,1{]}^2$ admits a cyclic ordering $z_1,\dots ,z_n$ such thatHere is a proof of that geometric result. First consider a right triangle with hypotenuse endpoints $a,b$, hypotenuse length $c$, and right-angle vertex $v$. The quadratic path bound in a right triangle supplies a path from $a$ to $b$ through any finite set inside the triangle whose sum of squared edge lengths is at most $c^2$. For an empty set use the edge $ab$. For a single point $z$, use $a,z,b$: the triangle lies in the disk with diameter $ab$, so $\angle azb\ge \pi /2$ and the law of cosines gives $\Vert a-z{\Vert }^2+\Vert z-b{\Vert }^2\le c^2$.

For several points, draw the altitude from $v$ to the hypotenuse. It splits the triangle into two smaller right triangles with hypotenuses $av$ and $vb$. Their squared hypotenuse lengths sum to $c^2$ by the Pythagorean theorem. Repeated altitude subdivision eventually makes every cell small enough to contain at most one of the given distinct points: each child is similar to the original triangle and its diameter contracts by at most the fixed factor $\max (\Vert a-v\Vert ,\Vert b-v\Vert )/c<1$. Construct the base-case paths and combine them from the bottom of this finite subdivision tree. The two child paths concatenate at $v$. If $v$ is an auxiliary point, delete it by joining its two neighbours directly. Both neighbours lie in the right-angle sector of the parent triangle at $v$, so their angle at $v$ is at most $\pi /2$. The law of cosines now shows that this shortcut does not increase the sum of squared edge lengths. The combined path still has endpoints $a,b$ and squared cost at most $c^2$.

One can first put the given points in generic positions, avoiding subdivision boundaries, and then pass to a limit. There are only finitely many possible visiting orders, so a subsequence keeps the same order and its squared cost converges; this also handles coincident locations. No assumption about a minimum separation remains in the result.

Split the square along a diagonal $ab$. Apply the triangle result in each half, following one path from $a$ to $b$ and the other back to $a$. Each half contributes at most $\Vert a-b{\Vert }^2=2$. Remove either diagonal endpoint if it was auxiliary. At a square corner all neighbours lie in a sector of angle $\pi /2$, so the same squared-cost shortcut applies. This gives the asserted cyclic ordering with total squared cost at most four. It does not assert that the tour minimizing ordinary length itself has this squared-edge property. The Cauchy-Schwarz inequality applied to the tour supplied by the theorem provesThus the upper bound is deterministic, independently of the sampling law.

For the lower bound, put $R_i={\min }_{j\ne i}\Vert X_i-X_j\Vert$, the nearest neighbour distance at $X_i$. Both edges incident to $X_i$ in any closed tour have length at least $R_i$. Summing over vertices and then dividing by two gives $Y_n\ge {\sum }_i{R}_i$. Conditional on $X_i=x$, the union bound and the uniform distribution giveBoundary clipping only makes the ball smaller. Apply the layer-cake representation of the expected value, restricting the integral to $0\le t\le 1/(2\sqrt{n})$:Linearity of the expected value now yieldsThe unspecified sampling phrase in the paper must therefore be understood as independent uniform sampling: an arbitrary distribution concentrated in a tiny region cannot satisfy this lower bound.

Let $X_s$ count the copies of the complete graph $K_s$ in the binomial random graph $G_{n,p}$. By linearity of the expected value and independence of distinct edges,Throughout, $n$ is sufficiently large that $k=\lfloor \log \log n\rfloor \ge 2$. We have $k\to \infty$, $k=O(\log \log n)$ and $k\log k=o(\log n)$.

First, since $p=n^{-2/k}$,The first moment method proves $\mathbb{P}(X_{k+1}>0)=o(1)$.

For the existence of a $k$-clique, put $X=X_k$ and $\mu =\mathbb{E}X$. Since $p^{\left(\genfrac{}{}{0ex}{}{k}{2}\right)}=n^{-(k-1)}$,This alone is insufficient: we must control the variance via the overlap formula for the variance of a clique count. Let $I_S$ indicate that a given $k$-set $S$ spans a clique. If $|S\cap T|=j$, the two copies share $\left(\genfrac{}{}{0ex}{}{j}{2}\right)$ edges, soFor $j=0,1$ their edge sets are disjoint and the covariance vanishes, even when they share one vertex. Counting ordered pairs of $k$-sets gives the exact formulaThis includes the diagonal case $j=k$.

To bound the overlap factor, regard it as the probability that a uniformly chosen $k$-set $T$ intersects a fixed $S$ in exactly $j$ vertices. A union bound over the possible specified $j$-subsets of $S$ yieldsHere $(a{)}_j=a(a-1)\cdots (a-j+1)$ is a falling factorial, and the last bound holds for $n\ge 2k$. ThereforeFor $2\le j\le k$, the concave quadratic $j(k-j+1)$ is minimized at an endpoint. Its endpoint values are $2(k-1)$ and $k$, both at least $k$. Thus every power of $n$ in the last sum is at most $n^{-1}$, andbecause the logarithm of its numerator is $O(k\log k)=o(\log n)$. The second moment method, or Chebyshev's inequality directly, gives $\mathbb{P}(X=0)=o(1)$. Finally, every larger clique contains a $(k+1)$-clique, so the absence of $K_{k+1}$ excludes all larger sizes. The clique number is exactly $k$ with high probability: