Abstract
Let ξ p+11,...,ξ p+1n , p⩽1 n⩽p+2 be r.v.’s possessing the following properties: E(ξ p+1i ) exists for 1⩾i⩾n and
Further let the covariance matrix of (ξ p+11,...,ξ p+1n ) exist and be denoted by U = (u ij )\( _{1n}^{1n} \). Here, the x ji, 1⩾j⩾p, 1⩾i⩾n are given real numbers and the βi0⩾i⩾p, as well as the u ij, 1⩾i, j⩾n, real parameters. The βi satisfy -<∞βi<∞ and the u ij need only satisfy the trivial restriction that U be positive semi-definite. To be more precise, we should denote the right side of (1.1) by E(ξ p+1i;β0,...,βp) or even by E(ξ p+1i;β0,...,βp, uij, 1⩾i,j⩾n) but the abbreviated notation should cause no misunderstanding. Our task is to construct unbiased estimates for each βi0⩾i⩾p. In order to bring this problem into the general framework of V.1, we let the sample space be given by \( ({R_n},{\mathfrak{B}_n}) \) and the set of joint distributions of (ξ p+11,...,ξ p+1n ) be so restricted by the parameters β0,...,βp, uij,1⩾i,j⩾n that (1.1) holds and (uij)\( _{1n}^{1n} \) n is positive semi-definite. To obtain the estimates we make use of Gauss’ method of least squares, which is closely connected with the MLP.
This is a preview of subscription content, log in via an institution.
Buying options
Tax calculation will be finalised at checkout
Purchases are for personal use only
Learn about institutional subscriptionsPreview
Unable to display preview. Download preview PDF.
References
The bj provided they exist, will in general also depend on the xq, xp. Since, however, we view these n-tuples here as given, we suppress this dependence.
From the extensive literature on the method of least squares we indicate only: A. C. Aitken, Proc. Roy. Soc. Edinburgh Sect. A, 55, 42–48 (1935), B.J. van Ijzeren, Statistica Rijswijk 8, 21–45 (1954), O. Kempthorne.
See B.J. van Ijzeren, l.c.2.
See A. N. Kolmogorov, Uspehi. Mat. Nauk 1, 57–70 (1946).
A. Markov, Wahrscheinlichkeitsrechnung, 2nd ed., Leipzig-Berlin 1912. Also see F.N. David and J. Neyman, l.c. V.1 and L. Schmetterer, l.c. V.5, second paper listed.
It should cause no difficulty when the symbol 0 denotes both the (p + 1)-dimensional null-vector and zero itself.
See H. Scheffe, l.c. III72
For clarity, we now write S(xp+x) instead of S.
The r.v. M (p+1) as well as M(p+1) and M(p+1)/M1 (p+1) are undefined on a set of probability zero, i.e., on the set where the denominator in the definition vanishes.
This distribution was first found by J. Wishart, Biometrika 20A, 32–52 (1928). The induction proof here is due essentially to P. L. Hsu, Proc. Cambridge Philos. Soc. 35, 336–338 (1939).
Note that here, as was not the case in 1 and 2, the symbols xi and ξidenote k-dimensional vectors.
C.R. Rao, Sankhya 9, 343–366 (1949).
The matrix \( ({W_{ij}})_{1k}^{1k} \) exists iff \( |{w_{ij}}|_{1k}^{1k} \ne 0. \). The matrix \( ({w_{ij}})_{1k}^{1k} \) is positive-semi-definite.
This distribution was first discovered by H. Hotelling, Ann. Math. Statist. 2, 360–378 (1931). Also see H. Hotelling, Proceedings of 2nd Berkeley Symposium on Mathematical Statistics and Probability 1951, pp. 23–41 University of California Press, Berkeley and Los Angeles.
A comparison with the F-distribution shows that \( \frac{{n - k}}{{\left( {n - 1} \right)k}}{T^2} \) possesses an F-distribution with (k, n-k) degrees of freedom.
\( ({d_{ij}})_{1k}^{1k} \) denotes t n e inverse of the covariance matrix D-1.
R.A. Fisher, Ann. Eugenics 7, 179–188 (1936); P.C. Mahalanobis, Proc. Nat. Inst. Sci. India 2, 49–55 (1936).
We refer only to P. C. Mahalanobis, Sankhya 9,237–239 (1949) and papers by C. R. Rao, such as Biometrika 35, 58–79 (1948); Sankhya, l.c.12; Sankhya 10, 257–268 (1950).
Communicated by R. Borges.
The notation here is so chosen that Xj is a k-dimensional vector but xj is the real number \( \frac{1}{{{n_1}}}\sum\limits_{i = 1}^{{n_1}} {{x_{ji}}} \) and similarly for the yji.
SeeC.R. Rao, l.c.12.
That is the regression function of ξp+1 w.r.t.(ξ1,...,ξp), say.
G.U. Yule, Proc. Roy. Soc. London Ser. A, 79, 182–193 (1907).
It would perhaps be more consistent in the sense of the notation used in 1 and 2 to write bp+11.2...p instead of bp+11.2...p.
This terminology clearly points out the fact that one can calculate these quantities from a sample.
See M. S. Bartlett, Proc. Roy. Soc. Edinburgh Sect. A, 53, 260–283 (1932–1933). For a extension of this method see R.A. Wijsman, Ann. Math. Statist. 28, 415–422 (1957) and A.M. Kshirsagar, Ann. Math. Statist. 30, 239–241 (1959).
The symbol (ui.k...1) stands for the (p-K+1)-dimensional r.v. (up+1.k...1,...,uk+1.k...1).
First found by V. Romanovskij, Bull. Acad. Sci. Leningrad (6) 20, 643–648 (1926) and K. Pearson, Proc. Roy. Soc. London Ser. A, 112,1–14 (1926).
It is not entirely correct to use the symbol K2 p+1 here, but this should cause no confusion.
This distribution was first found by R.A. Fisher, Proc. Roy. Soc. London Ser. A, 121, 654–673 (1928).
R.A. Fisher pointed this out in this and other connections. See R.A. Fisher, Metron 3, 90–104 (1925).
R.A. Fisher, Biometrika 10, 507–521 (1915).
A.T. James, Ann. Math. Statist. 25, 40–75 (1954); A.G. Constantine and A.T. James, Ann. Math. Statist. 29, 1146–1166 (1958); A.T. James, Ann. Math. Statist. 31, 151–158 (1960) and A.T. James, Ann. Math. Statist. 32, 874–882 (1961), and others.
Author information
Authors and Affiliations
Rights and permissions
Copyright information
© 1974 Springer-Verlag Berlin · Heidelberg
About this chapter
Cite this chapter
Schmetterer, L. (1974). Theory of Regression and the Sampling Theory of Multidimensional Normal Distributions. In: Introduction to Mathematical Statistics. Die Grundlehren der mathematischen Wissenschaften, vol 202. Springer, Berlin, Heidelberg. https://doi.org/10.1007/978-3-642-65542-5_8
Download citation
DOI: https://doi.org/10.1007/978-3-642-65542-5_8
Publisher Name: Springer, Berlin, Heidelberg
Print ISBN: 978-3-642-65544-9
Online ISBN: 978-3-642-65542-5
eBook Packages: Springer Book Archive