Asymptotic distribution of $\mathbb E_{\hat{P}_n}[Z]^T\operatorname{Cov}_{\hat{P}_n}[Z]^{-1}\mathbb E_{\hat{P}_n}[Z]$

Under very general conditions on the random $$p$$-dimensional vector $$Z$$, what can be said about the asymptotic distribution of the (random) scalar quantity $$R_n := \mathbb E_{\hat{P}_n}[Z]^T\operatorname{Cov}_{\hat{P}_n}[Z]^{-1}\mathbb E_{\hat{P}_n}[Z]$$ ?

Here, $$\mathbb E_{\hat{P}_n}[Z] = (1/n)\sum_{i=1}^nz_i \in \mathbb R^p$$ is the empirical mean of $$Z$$ from an i.i.d sample $$z_1,\ldots,z_n$$, and $$\operatorname{Cov}_{\hat{P}_n}[Z] \in \mathbb R^{p \times p}$$ is the empirical covariance matrix.

Notes

My ultimate goal is to understand the rate of growth of $$R_n$$ as a function of $$n$$.

Observations

My wild guess is that $$R_n$$ should be "concentrated" around $$\mu^T\Sigma^{-1}\mu$$ where $$\mu \in \mathbb R^p$$ is the mean of $$Z$$ and $$\Sigma \in \mathbb R^{p \times p}$$ is its covariance matrix.

If $$\mu\ne0$$, then the distribution of $$R_n$$ is asymptotically normal with asymptotic mean $$\mu^T\Sigma^{-1}\mu$$ and an explicit asymptotic variance $$\tilde\sigma^2/n$$; see e.g. Theorem 3.9, page 1018, where a bound on the rate of convergence is also given. More specifically, $$$$\tilde\sigma^2=EL(V)^2=E(2\xi-\xi^2+\mu^T\Sigma^{-1}\mu)^2,$$$$ where $$$$V:=(Y-EY,(Y-EY)(Y-EY)^T-I),\quad Y:=\Sigma^{-1/2}z_1,$$$$ $$$$L(x_1,x_2):=2x_1^T\,EY-EY^T\,x_2\,EY,$$$$ $$$$\xi:=EY^T\,(Y-EY)=\mu^T\Sigma^{-1}(z_1-\mu).$$$$
If $$\mu=0$$, then the distribution of properly normalized $$R_n$$ is asymptotically chi-squared; see e.g. Theorem 3, page 48.
• Two short questions: (1) In Theorem 3.9 what's an explict formula for $\tilde{\sigma}$ (it's supposed to be defined in 2.21, but I kinda lost tract of the definition of $L(V)$ relative to Hotelling's statistic under study in Theorem 3.9. (2) Can the works in this paper be applied to get asymptotic distributions of $M$-estimators ? Thanks in advance. – dohmatob Oct 23 '18 at 23:06
• (1) I have now reproduced the explicit expression for $\tilde\sigma$. (2) Results of the first referenced paper do apply to $M$-estimators, albeit somewhat indirectly; see the comment right after Remark 7.3 in projecteuclid.org/euclid.ejs/1491897618 . – Iosif Pinelis Oct 24 '18 at 15:21
• Thanks for the generous update. OK, I see . However, concerning $M$-estimators, this extension is not completely trivial though. IMO, it deserves a separate treatment (post, manuscript etc.). I don't master enough statistics to do this myself without screwing certain things badly. I can make this into a separate question, and maybe you could roughly sketch the main steps ? Thanks. – dohmatob Oct 24 '18 at 16:37
• I thought the comment right after Remark 7.3 in the MLE paper should be enough. I guess I can try to state the result for $M$-estimators formally. However, I think this would rather be a separate question. – Iosif Pinelis Oct 24 '18 at 19:55
• If $\Sigma$ is not full rank, it just means that the underlying linear space was wrongly chosen: it should be the column space (say $V$) of $\Sigma$, preferably with a good basis of it, say an orthonormal eigenbasis or somewhat close to it. Then the matrix of the linear transformation $x\mapsto\Sigma x$ in that good basis will be somewhat close to a diagonal matrix, or even to the identity matrix after appropriate re-scaling. So, if $\mu\in V$, then we will have $\mu^T\Sigma^{-1}\mu\in[0,\infty)$. If $\mu\notin V$, then $\mu^T\Sigma^{-1}\mu$ should naturally be defined as $\infty$. – Iosif Pinelis Nov 14 '18 at 14:52