Abstract
This paper investigates the hypothesis test of the parametric component in partial functional linear quantile regression model in which the dependent variable is related to both a vector of finite length and a function-valued random variable as predictor variables. A quantile rank score test based on functional principal component analysis is developed. Under mild conditions, we establish the consistency of the proposed test statistic, and show that the proposed test can detect Pitman local alternatives converging to the null hypothesis at the usual parametric rate. A simulation study shows that the proposed test procedure has good size and power with finite sample sizes. Finally, an illustrative example is given through fitting the Berkeley growth data and testing the effect of gender on the height of kids.
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Acknowledgements
Du’s work is supported by the National Natural Science Foundation of China (11971045), the Science and Technology Project of Beijing Municipal Education Commission (KM201910005015), Young Talent program of Beijing Municipal Commission of Education (CIT&TCD201904021), and China Postdoctoral Science Foundation Funded Project (2019M653502). Zhang’s work is supported by the National Natural Science Foundation of China (11771032).
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Appendix 1
Appendix 1
In order to provide the proofs of the theorems, we first define some notations and give some preliminary results.
Write \({\hat{V}}_k(g)=\sum _{j=1}^m\frac{\langle {\hat{C}}_{z_kX},{\hat{v}}_j\rangle \langle {\hat{v}}_j,g\rangle }{{\hat{\lambda }}_j}\), \(V_k(g)=\sum _{j=1}^{\infty }\frac{\langle C_{z_kX},v_j\rangle \langle v_j,g\rangle }{\lambda _j}\) for \(g\in L^2[0,1]\), \(\hat{\varvec{\Delta }}={\hat{C}}_{\varvec{z}}-\{{\hat{V}}_k({\hat{C}}_{z_{l}X})\}_{k,l=1,\ldots ,p}\). Then it is easy to show that \(\varvec{\Delta }\) defined in condition C6 can be expressed as \(\varvec{\Delta }=C_{\varvec{z}}-\{{V_k(C_{z_{l}X})}\}_{k,l=1,\ldots ,p}\), and that \(\hat{\varvec{\Delta }}=\frac{1}{n}\varvec{Z}^T(\varvec{I}-\varvec{P})\varvec{Z}=\frac{1}{n}{\varvec{Z}^*}^T{\varvec{Z}^*}\) if \({\hat{\lambda }}_1>\cdots>{\hat{\lambda }}_n>0\) holds. Let \(\varvec{S}^*_n=\frac{1}{\sqrt{n}}\sum _{i=1}^{n}\varvec{z}_i^*\psi _\tau ({\epsilon }_i)\), \(R_i=\int _{0}^{1}\gamma _0(t)X_i(t)dt-\varvec{U}_i^T\varvec{\gamma }_0\), \(\delta =n^{-\frac{2b-1}{2(a+2b)}}\).
The following Lemma 1, which is from He and Shao (2000, Lemma 3.2), is presented here for easy reference.
Lemma 1
Let \(\{v_i(t), t\in R^q\},\;1\le i\le n\), be independent \(R^d\)-valued random variables with \(E\{v_i(t)\}=0\) for all t. Assume that there exist \(r_1>0\) and \(r_2>0\) such that for every \(s\in R^q\), \(0<\delta ^*\le 1\), \(1\le i\le n\)
Let
and
Then
for every \(r_3\ge 0\).
Lemma 2
Let \(u_i(\varvec{\gamma },\varvec{\gamma }_0)=\psi _\tau (Y_i-\varvec{U}_i^T\varvec{\gamma })\varvec{z}_i^*-\psi _\tau (Y_i-\varvec{U}_i^T\varvec{\gamma }_0)\varvec{z}_i^* -E\psi _\tau (Y_i-\varvec{U}_i^T\varvec{\gamma })\varvec{z}_i^*+E\psi _\tau (Y_i-\varvec{U}_i^T\varvec{\gamma }_0)\varvec{z}_i^*\). Under the regularity conditions of Theorem 1, for any \(L>0\), it has
Proof
We prove this lemma using Lemma 1. First, we verify the condition in (13), that is, there exist \(\nu\) and \(r>0\) such that
Noting that \(I(\epsilon \le \cdot )\) and \(F(\cdot )\) are monotone increasing functions, \(\psi _\tau (u)=\tau - I(u<0)\) is a step function with a jump at \(u=0\), \(\Vert \varvec{z}_i^*\Vert =O_p(1)\) and \(\Vert \varvec{U}_i\Vert =O_p(1)\), a simple calculation yields
which implies (15).
Second, it holds that
thus, we have
which implies
Moreover, we can get
Invoking Lemma 1 and Condition C2(b), we have
This completes the proof of Lemma 2. \(\square\)
Lemma 3
Under the regularity conditions of Theorem 1, for any \(L>0\), it has
Proof
Since \(E\psi _\tau (\epsilon )=\tau -F(0)=0\), where F is the conditional distribution function of \(\epsilon\). Using the Taylor expansion, we have
The remainder is \(o\big (1/\sqrt{n}\big )\) because \(\Vert \varvec{\gamma }-\varvec{\gamma }_0\Vert \le L\delta\) and \(b>\frac{a}{2}+1\). Then, we can obtain that
where \(b=f(0)\) and \(c=f'(0)\). Note that \(\varvec{U}\) and \(\varvec{Z}^*\) are orthogonal to each other, that is, \(\varvec{U}^T\varvec{Z}^*=0\). Moreover,
Thus completes the proof of Lemma 3. \(\square\)
Proof of Theorem 1
First, note that
For \(\text {A}_1\), by condition C1 and the Hölder inequality, it follows
As for \(\text {A}_2\), due to
and
one has
Taking these together, we have
According to Theorem 3.2 of Kato (2012), we can get the fact that \(|\hat{\varvec{\gamma }}-\varvec{\gamma }_0|=O_p(\delta )\). Therefore, we can derive from Lemmas 2 and 3 that
Moreover, invoking Eq. (17) and following the similar arguments used for Eq. (18), a simple calculation yields
Combining (18) and (19), we have
In addition, it’s easy to see that
By the central limit theorem, we have
Then, Theorem 1 follows immediately from (20) and (21). \(\square\)
Lemma 4
Define
where \(\hat{\varvec{\gamma }}=\text {argmin}_{{\varvec{\gamma }}\in R^m}\sum _{i=1}^{n}\rho _\tau (Y_i-\varvec{U}_i\varvec{\gamma }-\varvec{z}_i^T\frac{\varvec{\beta }_0}{\sqrt{n}})\). Then under \({\widetilde{H}}_A\) and Conditions C1–C7, we have
Proof
Since \(E\psi _\tau (\epsilon )=\tau -F(0)\) and
Then, invoking (17) and the orthogonality between \(\varvec{U}\) and \(\varvec{Z}^*\), we have
This completes the proof of Lemma 4. \(\square\)
Proof of Theorem 2
According to Lemma 1 in Yu et al. (2016), we have
Note that \({\varvec{{S}}}_n^*\) is the summation of independent entries. It follows from the Lindberg-Feller central limit theorem that
The combination of (23), (24), Lemma 4 and the definition of non-central Chi-squared distribution allow us to finish the proof of Theorem 2. \(\square\)
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Yu, P., Du, J. & Zhang, Z. Testing linearity in partial functional linear quantile regression model based on regression rank scores. J. Korean Stat. Soc. 50, 214–232 (2021). https://doi.org/10.1007/s42952-020-00070-9
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DOI: https://doi.org/10.1007/s42952-020-00070-9
Keywords
- Functional data analysis
- Functional linear quantile regression
- Functional principal component analysis
- Rank score test