A General Class of Estimators of Population Median Using Two Auxiliary Variables in Double Sampling by Florentin Smarandache

Mohammad Khoshnevisan School of Accounting and Finance Griffith University, Australia

Housila P. Singh School of Studies in Statistics, Vikram University Ujjain - 456 010 (M. P.), India

Sarjinder Singh

Departament of Mathematics and Statistics University of Saskatchewan, Canada

Florentin Smarandache University of New Mexico, Gallup, USA

A General Class of Estimators of Population Median Using Two Auxiliary Variables in Double Sampling

Published in: M. Khoshnevisan, S. Saxena, H. P. Singh, S. Singh, F. Smarandache (Editors) RANDOMNESS AND OPTIMAL ESTIMATION IN DATA SAMPLING (second edition) American Research Press, Rehoboth, USA, 2002 ISBN: 1-931233-68-3 pp. 26 - 43

Abstract: In this paper we have suggested two classes of estimators for population median MY of the study character Y using information on two auxiliary characters X and Z in double sampling. It has been shown that the suggested classes of estimators are more efficient than the one suggested by Singh et al (2001). Estimators based on estimated optimum values have been also considered with their properties. The optimum values of the first phase and second phase sample sizes are also obtained for the fixed cost of survey. Keywords: Median estimation, Chain ratio and regression estimators, Study variate, Auxiliary variate, Classes of estimators, Mean squared errors, Cost, Double sampling. 2000 MSC: 60E99

1. INTRODUCTION In survey sampling, statisticians often come across the study of variables which have highly skewed distributions, such as income, expenditure etc. In such situations, the estimation of median deserves special attention. Kuk and Mak (1989) are the first to introduce the estimation of population median of the study variate Y using auxiliary information in survey sampling. Francisco and Fuller (1991) have also considered the problem of estimation of the median as part of the estimation of a finite population distribution function. Later Singh et al (2001) have dealt extensively with the problem of estimation of median using auxiliary information on an auxiliary variate in two phase sampling. Consider a finite population U={1,2,…,i,...,N}. Let Y and X be the variable for study and auxiliary variable, taking values Yi and Xi respectively for the i-th unit. When the two variables are strongly related but no information is available on the population median MX of X, we seek to estimate the population median MY of Y from a sample Sm, obtained through a two-phase selection. Permitting simple random sampling without replacement (SRSWOR) design in each phase, the two-phase sampling scheme will be as follows: (i)

The first phase sample Sn(Sn⊂U) of fixed size n is drawn to observe only X in order to furnish an estimate of MX.

(ii)

Given Sn, the second phase sample Sm(Sm⊂Sn) of fixed size m is drawn to observe Y only.

Assuming that the median MX of the variable X is known, Kuk and Mak (1989) suggested a ratio estimator for the population median MY of Y as

M Mˆ 1 = Mˆ Y X Mˆ X

(1.1)

ˆ and Mˆ are the sample estimators of MY and MX respectively based on a sample Sm of size where M Y X m. Suppose that y(1), y(2), …, y(m) are the y values of sample units in ascending order. Further, let t be an integer such that Y(t) ≤ MY ≤Y(t+1) and let p=t/m be the proportion of Y, values in the sample that are less ˆ is a predictor of p, the than or equal to the median value MY, an unknown population parameter. If p ˆ can be written in terms of quantities as Qˆ ( pˆ ) where pˆ = 0.5 . Kuk and Mak sample median M Y Y (1989) define a matrix of proportions (Pij(x,y)) as Y ≤ MY P11(x,y) P12(x,y) P1⋅(x,y)

X ≤ MX X > MX Total

Y > MY P21(x,y) P22(x,y) P2⋅(x,y)

Total P⋅1(x,y) P⋅2(x,y) 1

and a position estimator of My given by ( p) Mˆ Y = Qˆ Y ( pˆ Y )

where

pˆ Y =

(1.2)

1  mx pˆ 11 ( x, y ) (m − mx ) pˆ 12 ( x, y )    + m  pˆ ⋅1 ( x, y ) pˆ ⋅2 ( x, y ) 

 m pˆ ( x, y ) + (m − mx ) pˆ 12 ( x, y )  ≈ 2 x 11  m   ˆ ij ( x, y ) being the sample analogues of the Pij(x,y) obtained from the population and mx the number with p of units in Sm with X ≤ MX.

Let FYA ( y ) and FYB ( y ) denote the proportion of units in the sample Sm with X ≤ MX, and X>MX, respectively that have Y values less than or equal to y. Then for estimating MY, Kuk and Mak (1989) suggested the 'stratification estimator' as

{

}

~ ( y) ( St ) Mˆ Y = inf y : FY ≥ 0.5

[

1 ~ ( y) ~ ( y) FYA + FYB where FˆY ( y ) ≅ 2

(1.3)

]

It is to be noted that the estimators defined in (1.1), (1.2) and (1.3) are based on prior knowledge of the median MX of the auxiliary character X. In many situations of practical importance the population median MX of X may not be known. This led Singh et al (2001) to discuss the problem of estimating the population median MY in double sampling and suggested an analogous ratio estimator as

Mˆ 1X ˆ ˆ M 1d = M Y Mˆ X

(1.4)

ˆ 1 is sample median based on first phase sample Sn. where M X Sometimes even if MX is unknown, information on a second auxiliary variable Z, closely related to X but compared X remotely related to Y, is available on all units of the population. This type of situation has been briefly discussed by, among others, Chand (1975), Kiregyera (1980, 84), Srivenkataramana and Tracy (1989), Sahoo and Sahoo (1993) and Singh (1993). Let MZ be the known population median of Z. Defining

ˆ 1  Mˆ   Mˆ   Mˆ 1   Mˆ  M  e0 =  Y − 1, e1 =  X − 1, e2 =  X − 1, e3  Z − 1 and e 4 =  Z − 1  MY  MX   MX   MZ   MZ 

ˆ and Mˆ are the sample median estimators based such that E(ek)≅0 and ek<1 for k=0,1,2,3; where M 2 2 on second phase sample Sm and first phase sample Sn. Let us define the following two new matrices as 1

X ≤ MX X > MX Total

Z ≤ MZ P11(x,z) P12(x,z) P1⋅(x,z)

Z > MZ P21(x,z) P22(x,z) P2⋅(x,z)

Total P⋅1(x,z) P⋅2(x,z) 1

Y ≤ MY Y > MY Total

Z ≤ MZ P11(y,z) P12(y,z) P1⋅(y,z)

Z > MZ P21(y,z) P22(y,z) P2⋅(y,z)

Total P⋅1(y,z) P⋅2(y,z) 1

and

Using results given in the Appendix-1, to the first order of approximation, we have N-m E(e02) =  N  (4m)-1{MYfY(MY)}-2,   N-m E(e12) =  N  (4m)-1{MXfX(MX)}-2,   N-n E(e22) =  N  (4n)-1{MXfX(MX)}-2,   N-m E(e32) =  N  (4m)-1{MZfZ(MZ)}-2,   N-n E(e42) =  N  (4n)-1{MZfZ(MZ)}-2,   N-m E(e0e1) =  N  (4m)-1{4P11(x,y)-1}{MXMYfX(MX)fY(MY)}-1,   N-n E(e0e2) =  N  (4n)-1{4P11(x,y)-1}{MXMYfX(MX)fY(MY)}-1,   N-m E(e0e3) =  N  (4m)-1{4P11(y,z)-1}{MYMZfY(MY)fZ(MZ)}-1,   N-n E(e0e4) =  N  (4n)-1{4P11(y,z)-1}{MYMZfY(MY)fZ(MZ)}-1,   N-n E(e1e2) =  N  (4n)-1{MXfX(MX)}-2,   N-m E(e1e3) =  N  (4m)-1{4P11(x,z)-1}{MXMZfX(MX)fZ(MZ)}-1,  

N-n E(e1e4) =  N  (4n)-1{4P11(x,z)-1}{MXMZfX(MX)fZ(MZ)}-1,   N-n E(e2e3) =  N  (4n)-1{4P11(x,z)-1}{MXMZfX(MX)fZ(MZ)}-1,   N-n E(e2e4) =  N  (4n)-1{4P11(x,z)-1}{MXMZfX(MX)fZ(MZ)}-1,   N-n E(e3e4) =  N  (4n)-1(fZ(MZ)MZ)-2   where it is assumed that as N→∞ the distribution of the trivariate variable (X,Y,Z) approaches a continuous distribution with marginal densities fX(x), fY(y) and fZ(z) for X, Y and Z respectively. This assumption holds in particular under a superpopulation model framework, treating the values of (X, Y, Z) in the population as a realization of N independent observations from a continuous distribution. We also assume that fY(MY), fX(MX) and fZ(MZ) are positive.

ˆ is consistent and asymptotically normal (Gross, 1980) with Under these conditions, the sample median M Y mean MY and variance  N −m −1 −2  (4m ) { f Y (M Y )}  N  In this paper we have suggested a class of estimators for MY using information on two auxiliary variables X and Z in double sampling and analyzes its properties. 2. SUGGESTED CLASS OF ESTIMATORS Motivated by Srivastava (1971), we suggest a class of estimators of MY of Y as

{

}

(g ) (g ) g = Mˆ Y : Mˆ Y = M Y g (u , v )

where u =

Mˆ X ,v = Mˆ 1 X

(2.1)

Mˆ Z1 and g(u,v) is a function of u and v such that g(1,1)=1 and such that it satisfies Mˆ Z

the following conditions. 1.

Whatever be the samples (Sn and Sm) chosen, let (u,v) assume values in a closed convex subspace, P, of the two dimensional real space containing the point (1,1).

The function g(u,v) is continuous in P, such that g(1,1)=1.

The first and second order partial derivatives of g(u,v) exist and are also continuous in P.

Expanding g(u,v) about the point (1,1) in a second order Taylor's series and taking expectations, it is found that

(

)

( ) E Mˆ Y g = M Y + 0(n −1 ) so the bias is of order n−1. Using a first order Taylor's series expansion around the point (1,1) and noting that g(1,1)=1, we have

( )

(g ) Mˆ Y ≅ M Y [1 + e0 + (e1 − e2 )g1 (1,1) + e4 g 2 (1,1) + 0 n −1 ]

(M ( ) − M ) ≅ M [e + (e g

− e2 )g 1 (1,1) + e4 g 2 (1,1)] (2.2)

where g1(1,1) and g2(1,1) denote first order partial derivatives of g(u,v) with respect to u and v respectively around the point (1,1).

ˆ ( g ) to the first degree Squaring both sides in (2.2) and then taking expectations, we get the variance of M Y of approximation, as

(

)

(g ) Var Mˆ Y =

where

4( f Y (M Y ))

 1 1   1 1  1 1    m − N  +  m − n  A +  n − N  B ,       

(2.3)

 M f (M )    M f (M )  A =  Y Y Y  g1 (1,1)  Y Y Y  g1 (1,1) + 2(4 P11 ( x, y ) − 1)  M X f X (M X )   M X f X (M X )  

(2.4)

 M f (M )    M f (M )  B =  Y Y Y  g Z (1,1) Y Y Y  g 2 (1,1) + 2(4 P11 ( y, z ) − 1)  M Z f Z (M Z )   M Z f Z (M Z )  

(2.5)

ˆ ( g ) in (2.3) is minimized for The variance of M Y

 M f (M X )  g1 (1,1) = − X X (4 P11 ( x, y ) − 1)  M Y f Y (M Y )   M f (M Z )  g 2 (1,1) = − Z Z (4 P11 ( y, z ) − 1)  M Y f Y (M Y )  Thus the resulting (minimum) variance of M Y

(

)

min. Var Mˆ Y ( g ) =

1 4( f Y (M Y ))2

(g )

(2.6)

is given by

 1 1   1 1   2 1 1   m − N  −  m − n (4 P11 ( x, y ) − 1) −  n − N (4 P11 ( y, z ) − 1) (2.7)

Now, we proved the following theorem. Theorem 2.1 - Up to terms of order n-1,

(

) 4( f

ˆ g ≥ Var M Y

1 2 y (M Y )

)

 1 1   1 1  2 2 1 1   m − N  −  m − n (4 P11 ( x, y ) − 1) −  n − N (4 P11 ( y, z ) − 1) 

with equality holding if

 M f (M )  g1 (1,1) = − x x x (4 P11 ( x, y ) − 1)  M Y f Y (M Y )   M f (M z )  g 2 (1,1) = − z z (4 P11 ( y, z ) − 1)  M Y f Y (M Y )  ˆ ( g ) at (2.1) is the variance of the linear It is interesting to note that the lower bound of the variance of M y regression estimator

(

)

(

(l ) Mˆ Y = Mˆ Y + dˆ1 Mˆ 1X − Mˆ X + dˆ 2 M Z − Mˆ Z1

)

(2.8)

where

( ( fˆ (Mˆ fˆ (Mˆ

) ) ) (4 pˆ )

fˆ Mˆ x (4 pˆ 11 ( x, y ) − 1), dˆ1 = X fˆ Mˆ dˆ 2 =

( y, z ) − 1),

ˆ 11 ( x, y ) and pˆ 11 ( y, z ) being the sample analogues of the p11 ( x, y ) and p11 ( y, z ) respectively with p and

( )

fˆY Mˆ Y , fˆX (M X ) and fˆZ (M Z ) can be obtained by following Silverman (1986).

Any parametric function g(u,v) satisfying the conditions (1), (2) and (3) can generate an asymptotically acceptable estimator. The class of such estimators are large. The following simple functions g(u,v) give even estimators of the class

g (1) (u , v ) = u α v β , g ( 2 ) (u , v ) =

1 + α (u − 1) , 1 − β (v − 1)

g (3 ) (u, v ) = 1 + α (u − 1) + β (v − 1), g (4 ) (u , v ) = {1 − α (u − 1) − β (v − 1)} g (5 ) (u , v ) = w1u α + w2 v β , w1 + w2 = 1 g (6 ) (u , v ) = αu + (1 − α )v β , g (7 ) (u , v ) = exp{α (u − 1) + β (v − 1)}

−1

ˆ ( g ) = Mˆ g (i ) (u , v ), (i = 1 to7 ) . It is Let the seven estimators generated by g(i)(u,v) be denoted by M Yi Y easily seen that the optimum values of the parameters α,β,wi(i-1,2) are given by the right hand sides of (2.6). 3. A WIDER CLASS OF ESTIMATORS The class of estimators (2.1) does not include the estimator

(

)

(

)

Mˆ Yd = Mˆ Y + d1 Mˆ 1X − M X + d 2 M Z − Mˆ Z1 , (d1 , d 2 ) being constants.

However, it is easily shown that if we consider a class of estimators wider than (2.1), defined by

(

(G ) Mˆ Y = G1 Mˆ Y , u, v

)

(3.1)

Mˆ Y , u and v such that G (M Y ,1,1) = M Y and G1 (M Y ,1,1) = 1 . G1 (M Y ,1,1) denoting the first partial derivative of G(⋅) with respect to Mˆ Y .

of MY, where G(⋅) is a function of

Proceeding as in Section 2 it is easily seen that the bias of

ˆ (G ) is given by terms, the variance of M Y

(

)

ˆ (G ) = Var M Y

4( f Y (M Y ))

(G ) Mˆ Y is of the order n−1 and up to this order of

 1 1   1 1  f Y (M Y )   [  −  +  −   m N   m n  M X f X (M X ) 

 f Y (M Y )   G2 (M Y ,1,1) + 2(4 P11 ( x, y ) − 1) G2 (M Y ,1,1)  M X f X (M X )    1 1  f Y (M Y ) + −   n N  f Z (M Z )M Z

 f Y (M Y )   G3 (M Y ,1,1) + 2(4 P11 ( y, z ) − 1) ]   M Z f Z (M Z )   (3.2)

where G2(MY1,1) and G3(MY1,1) denote the first partial derivatives of u and v respectively around the point (MY,(1,1).

ˆ (G ) is minimized for The variance of M Y  M f (M X )  (4 P11 ( x, y ) − 1) G2 (M Y ,1,1) = − X X  f Y (M Y )   M f (M Z )  (4 P11 ( y, z ) − 1) G3 (M Y ,1,1) = − Z Z  f Y (M Y )  Substitution of (3.3) in (3.2) yields the minimum variance of

(

)

(3.3)

(G ) Mˆ Y as

1  1 1   1 1 1 1  2 2 [ −  −  − (4 P11 ( x, y ) − 1) −  − (4 P11 ( y, z ) − 1) ] 2  4( f Y (M Y ))  m N   m n  n N  ˆ (g) = min.Var M

ˆ (G ) = min. Var M Y

(

)

(3.4) Thus we established the following theorem. Theorem 3.1 - Up to terms of order n-1,

(

)

ˆ (G ) ≥ Var M Y

1 2 4( f Y (M Y ))

 1 1   1 1  1 1  2 2  m − N  −  m − n (4 P11 ( x, y ) − 1) −  n − N (4 P11 ( y, z ) − 1)        

with equality holding if

 f (M X )M X  (4 P11 ( x, y ) − 1) G2 (M Y ,1,1) = − x  f Y (M Y )   M f (M Z )  (4 P11 ( y, z ) − 1) G3 (M Y ,1,1) = − Z Z ( ) f M Y Y   ˆ (G ) reduces to If the information on second auxiliary variable z is not used, then the class of estimators M Y the class of estimators of MY as

(

(H ) Mˆ Y = H Mˆ Y , u

(

)

(

)

(3.5)

)

ˆ , u is a function of Mˆ , u such that H (M ,1) = M and H (M ,1) = 1, where H M Y Y Y Y 1 Y ∂H (⋅)  ˆ ( H ) is reported by Singh et al (2001). . The estimator M  Y ∂Mˆ Y  ( M ,1) Y

H 1 (M Y ,1) =

The minimum variance of

(

ˆ min.Var M Y

(H )

(H ) Mˆ Y to the first degree of approximation is given by

) = 4( f

1 2 Y (M Y ))

 1 1   1 1  2  m − N  −  m − n (4 P11 ( x, y ) − 1)      

(3.6)

From (3.4) and (3.6) we have

(

)

(

)

1 ˆ ( H ) − min.Var M ˆ (G ) =  1 − 1  (4 P11 ( y, z ) − 1)2 minVar M Y Y 2  n N  4( f Y (M Y )) ˆ which is always positive. Thus the proposed class of estimators M Y

(G )

(3.7)

is more efficient than the estimator

(H ) Mˆ Y considered by Singh et al (2001).

4. ESTIMATOR BASED ON ESTIMATED OPTIMUM VALUES We denote

α1 =

M X f X (M X ) (4 P11 ( x, y ) − 1) M Y f Y (M Y )

M f (M Z ) (4 P11 ( y, z ) − 1) α2 = Z Z M Y f Y (M Y )

(4.1)

In practice the optimum values of g1(1,1)(=-α1) and g2(1,1)(=-α2) are not known. Then we use to find out their sample estimates from the data at hand. Estimators of optimum value of g1(1,1) and g2(1,1) are given as

gˆ 1 (1,1) = −αˆ 1 gˆ 2 (1,1) = −αˆ 2

(4.2)

where

( ( ( (

) ) ) )

fˆX Mˆ X (4 pˆ 11 ( x, y ) − 1) ˆ Mˆ f Y Y Y ˆ Mˆ f Mˆ αˆ 2 = Z Z Z (4 p11 ( y, z ) − 1) Mˆ Y fˆY Mˆ Y αˆ 1 =

Mˆ X Mˆ

(4.3)

Now following the procedure discussed in Singh and Singh (19xx) and Srivastava and Jhajj (1983), we define the following class of estimators of MY (based on estimated optimum) as ( g *) = Mˆ Y g * (u , v, αˆ 1 , αˆ 2 ) Mˆ Y

(4.4)

where g*(⋅) is a function of (u , v, αˆ 1 , αˆ 2 ) such that

g * (1,1,α 1α 2 ) = 1 g1* (1,1,α 1 , α 2 ) =

∂g * (⋅) = −α 1 ∂u (1,1,α1 ,α 2 )

g 2* (1,1,α 1 , α 2 ) =

∂g * (⋅) = −α 2 ∂v (1,1,α1 ,α 2 )

g 3* (1,1,α 1 , α 2 ) =

∂g * (⋅) =0 ∂αˆ 1 (1,1,α ,α ) 1

g 4* (1,1,α 1 , α 2 ) =

∂g * (⋅) =0 ∂αˆ 2 (1,1,α ,α ) 1

and such that it satisfies the following conditions: 1.

2. 3.

Whatever be the samples (Sn and Sm) chosen, let u , v, αˆ 1αˆ 2 assume values in a closed convex subspace, S, of the four dimensional real space containing the point (1,1,α1,α2). The function g*(u,v, α1, α2) continuous in S. The first and second order partial derivatives of g * (u , v, αˆ 1 , αˆ 2 ) exst. and are also continuous in S.

Under the above conditions, it can be shown that

(

)

( )

( g *) E Mˆ Y = M Y + 0 n −1

and to the first degree of approximation, the variance of

(

( g *) is given by Mˆ Y

)

(

( g *) ˆ g Var Mˆ Y = min.Var M Y

(

)

(4.5)

)

ˆ ( g ) is given in (2.7). where min.Var M Y A wider class of estimators of MY based on estimated optimum values is defined by

(

(G* ) Mˆ Y = G * Mˆ Y , u , v, αˆ 1* , αˆ 2*

)

(4.6)

where

( ) ( ) Mˆ fˆ (Mˆ ) (4 pˆ = fˆ (Mˆ )

αˆ 1* =

Mˆ X fˆX Mˆ X (4 pˆ 11 ( x, y ) − 1) fˆ Mˆ Y

αˆ 2*

(4.7)

( y, z ) − 1)

are the estimates of

α 1* =

M x f X (M X ) (4 P11 ( x, y ) − 1) f Y (M Y )

M f (M Z ) (4 P11 ( y, z ) − 1) α = Z Z f Y (M Y ) * 2

(

)

ˆ , u , v, α , αˆ such that and G*(⋅) is a function of M Y 1 2

(

)

(

)

G * M Y ,1,1,α1* ,α 2* = M Y G1* M Y ,1,1,α 1* ,α 2* =

∂G * (⋅) ∂Mˆ Y (M

* * Y ,1,1,α1 ,α 2

)

(

)

∂G * (⋅) = −α 1* ∂u (M Y ,1,1,α1* ,α 2* )

(

)

∂G * (⋅) = −α 2* ∂v (M Y 1,1,∂1*,α 2* )

(

)

∂G * (⋅) ∂αˆ 1* (M

G2* M Y 1,1,α 1* ,α 2* =

G3* M Y 1,1, α 1* ,α 2* =

G4* M Y 1,1, α 1* ,α 2* =

* * Y ,1,1,α1 ,α 2

)

(4.8)

(

)

G5* M Y 1,1, α1* , α 2* =

∂G * (⋅) ∂αˆ 2* (M

* * Y ,1,1,α1 ,α 2

)

Under these conditions it can be easily shown that

(

)

( )

(G* ) E Mˆ Y = M Y + 0 n −1

ˆ (G*) is given by and to the first degree of approximation, the variance of M Y

(

)

(

(G ) G* Var Mˆ Y = min.Var Mˆ Y

where

(

)

(4.9)

)

G min.Var Mˆ Y is given in (3.4).

ˆ ( g *) and It is to be mentioned that a large number of estimators can be generated from the classes M Y (G * ) based on estimated optimum values. Mˆ Y

5. EFFICIENCY OF THE SUGGESTED CLASS OF ESTIMATORS FOR FIXED COST

ˆ , The appropriate estimator based on on single-phase sampling without using any auxiliary variable is M Y whose variance is given by

( )

1 1 1 Var Mˆ Y =  −  2  m N  4( f Y (M Y ))

(5.1)

In case when we do not use any auxiliary character then the cost function is of the form C0-mC1, where C0 and C1 are total cost and cost per unit of collecting information on the character Y. The optimum value of the variance for the fixed cost C0 is given by

  G 1  Opt.Var Mˆ Y = V0  −  C N  0  

( )

(5.2)

where

4( f Y (M Y ))

(5.3)

When we use one auxiliary character X then the cost function is given by

C 0 = Gm + C 2 n,

(5.4)

where C2 is the cost per unit of collecting information on the auxiliary character Z.

ˆ ( H ) is optimum, are The optimum sample sizes under (5.4) for which the minimum variance of M Y

m opt =

[

(V0 − V1 ) / C1 (V0 − V1 )C1 + V1C 2 ] C0

n opt =

[

C 0 V1 / C 2

(V0 − V1 )C1 +

V1C 2

(5.5)

]

where V1=V0(4P11(x,y)-1)2.

ˆ ( H ) in (3.6), we get Putting these optimum values of m and n in the minimum variance expression of M Y the optimum

(

)

(H ) as min.Var Mˆ Y

)] (

 (H ) Opt. min.Var Mˆ Y =  

[

(

(V0 − V1 )C1 +

V1C 2

)

−

V0   N 

(5.7)

Similarly, when we use an additional character Z then the cost function is given by

C 0 = C1 m + (C 2 + C 3 )n

(5.8)

where C3 is the cost per unit of collecting information on character Z. It is assumed that C1>C2>C3. The optimum values of m and n for fixed cost C0 which minimizes the

ˆ minimum variance of M Y

(g )

(orMˆ ) (2.7) (or (3.4)) are given by (G )

m opt =

n opt =

[

(V0 − V1 ) C1 (V0 − V1 )C1 + (C 2 + C3 )(V1 − V2 ) ] C0

(5.9)

(V1 − V2 ) C 2 + C3 (V0 − V1 )C1 + (C 2 + C3 )(V1 − V2 ) ] C0

(5.10)

where V2=V0(4P11(y,z)-1)2. The optimum variance of

(

)

(g ) (G ) corresponding to optimal two-phase sampling strategy is Mˆ Y orMˆ Y

[ (V0 − V1 )C1 + (C 2 + C 3 )(V1 − V2 )] 2 V2  (g ) (G ) ˆ ˆ Opt min.Var M Y or min.Var M Y = −  C0 N  

[

(

)

(

)]

(5.11) Assuming large N, the proposed two phase sampling strategy would be profitable over single phase sampling so long as

[Opt.Var(Mˆ )] > Opt.[min.Var(Mˆ ( ) )or min.Var(Mˆ ( ) )] g

i.e.

C 2 + C 3  V0 − V0 − V1  <  C1 V1 − V2  

(5.12)

When N is large, the proposed two phase sampling is more efficient than that Singh et al (2001) strategy if

[

(

)

(

)]

[

(

(g ) (G ) (H ) Opt min.Var Mˆ Y or min.Var Mˆ Y < Opt min.Var Mˆ Y

i.e.

)]

C 2 + C3 V1 < C1 V1 − V2

(5.13)

6. GENERALIZED CLASS OF ESTIMATORS We suggest a class of estimators of MY as

{

(

(F ) (F ) ℑ = Mˆ Y : Mˆ Y = F Mˆ Y , u, v, w

)}

(6.1)

ˆ / Mˆ ′ , v = Mˆ ′ / M , w = Mˆ / M and the function F(⋅) assumes a value in a where u = M X X Z Z Z Z bounded closed convex subset W⊂ℜ4, which contains the point (MY,1,1,1)=T and is such that

ˆ around the F(T)=MY⇒F1(T)=1, F1(T) denoting the first order partial derivative of F(⋅) with respect to M Y point T=(MY,1,1,1). Using a first order Taylor's series expansion around the point T, we get

(

)

(F ) Mˆ Y = F (T ) + Mˆ Y = M Y F1 (T ) + (u − 1) F2 (T ) + (v − 1) F3 (T ) + ( w − 1) F4 (T ) + 0(n −1 )

(6.2)

(

)

ˆ , u, v, w with respect to u, where F2(T), F3(T) and F4(T) denote the first order partial derivatives of F M Y v and w around the point T respectively. Under the assumption that F(T)=MY and F1(T)=1, we have the following theorem. Theorem 6.1. Any estimator in ℑ is asymptotically unbiased and normal. Proof: Following Kuk and Mak (1989), let PY, PX and PZ denote the proportion of Y, X and Z values respectively for which Y≤MY, X≤MX and Z≤MZ; then we have

Mˆ Y − M Y = Mˆ X − M X =

1 1 (1 − 2 PY ) + 0 p  n − 2 ,   2 f Y (M Y ) 1 1 (1 − 2 PX ) + 0 p  n − 2 ,   2 f X (M X )

Mˆ x′ − M X =

1 1 (1 − 2 PX ) + 0 p  n − 2    2 f X (M X )

Mˆ z − M Z =

1 1 (1 − 2 PZ ) + 0 p  n − 2    2 f Z (M Z )

and

Mˆ Z′ − M Z =

1 1 (1 − 2 PZ ) + 0 p  n − 2    2 f Z (M z )

Using these expressions in (6.2), we get the required results. Expression (6.2) can be rewritten as

(

)

(F ) Mˆ Y − M Y ≅ Mˆ Y − M Y + (u − 1)F2 (T ) + (v − 1) F3 (T ) + ( w − 1) F4 (T )

or (F ) Mˆ Y − M Y ≅ M Y e0 + (e1 − e2 )F2 (T ) + e4 F3 (T ) + e3 F4 (T )

(6.3)

ˆ ( F ) to the first degree Squaring both sides of (6.3) and then taking expectation, we get the variance of M Y of approximation, as

(

)

(F ) Var Mˆ Y =

1 2 4( f Y (M Y ))

 1 1   1 1 1 1    m − N  A1 +  m − n  A2 +  n − N  A3 ,       

where

  f (M )  2 2   f Y (M Y )  Y Y  F4 (T ) + 2(4 P11 ( y, z ) − 1)  F4 (T ) A1 = 1 +    M Z f Z (M Z )    M Z f Z (M Z )   f Y (M Y )  2   F2 (T ) + 2(4 P11 ( x, y ) − 1) F2 (T )   f Y (M Y )   M X f X (M X )    A2 =    ( ) M f M   X   X X  + 2(4 P11 ( x, z ) − 1 f Y (M Y )  F2 (T ) F4 (T )   M f (M )    Z   z Z  f Y (M Y )  2   F3 (T ) + 2(4 P11 ( y, z ) − 1) F3 (T )   f Y (M Y )   M Z f Z (M Z )    A3 =     f Y (M Y )   M Z f Z (M Z )    F3 (T ) F4 (T )  + 2    M Z f Z (M Z )  The

(

)

(F ) at (6.4) is minimized for Var Mˆ Y

(6.4)

F2 (T ) = −

[(4 P11 ( x, y ) − 1) − (4 P11 ( x, z ) − 1)(4 P11 ( y, z ) − 1)] M X f X (M X ) ⋅ ⋅ 2 f Y (M Y ) [1 − (4 P11 ( x, z ) − 1) ]

= − a 2 (say) (6.5)

F3 (T ) = −

(4 P11 ( x, z ) − 1)[(4 P11 ( x, y ) − 1) − (4 P11 ( y, z ) − 1)(4 P11 ( x, z ) − 1)] ⋅ M Z f Z (M Z ) ⋅ 2 f Y (M Y ) [1 − (4 P11 ( x, z ) − 1) ]

= −a 2 (say) F4 (T ) = −

[(4 P11 ( y, z ) − 1) − (4 P11 ( x, y ) − 1)(4 P11 ( x, z ) − 1)] M Z f Z (M Z ) ⋅ ⋅ 2 f Y (M Y ) [1 − (4 P11 ( x, z ) − 1) ]

= − a3 (say) Thus the resulting (minimum) variance of

(

(F ) minVar Mˆ Y

)

(F ) Mˆ Y is given by

2  1 1   D2 1 1      −  −  −  + (4 P11 ( x, y ) − 1)  1  m N   m n 1 − (4 P11 ( x, z ) − 1)2   = 2   4(f Y (M Y )) 1 1  2   −  − (4 P11 ( y, z ) − 1)   n N  D2 1  1 1 (G ) ˆ = min.Var M Y − −  2 2  m n  4( f Y (M Y )) 1 − 4 P11 ( x, z ) − 1

(

)

[

]

(6.6) where

D = [(4 P11 ( y, z ) − 1) − (4 P11 ( x, y ) − 1)(4 P11 ( x.z ) − 1)]

(

(6.7)

)

ˆ (G ) is given in (3.4) and min.Var M Y ˆ ( F ) is more efficient than the Expression (6.6) clearly indicates that the proposed class of estimators M Y

(

)

ˆ (G ) or Mˆ ( g ) and hence the class of estimators Mˆ ( H ) suggested by Singh et al class of estimator M Y Y Y

(2001) and the estimator

Mˆ Y at its optimum conditions.

The estimator based on estimated optimum values is defined by

{

(

F* ( F *) : Mˆ Y = F * Mˆ Y , u , v, w, aˆ1 , aˆ 2 , aˆ 3 p* = Mˆ Y

where

)}

(6.8)

aˆ1 =

[(4 pˆ 11 (x, y ) − 1) − (4 pˆ 11 (x, z ) − 1)(4 pˆ 11 ( y, z ) − 1)] Mˆ x fˆx (Mˆ x ) ⋅ 2 [1 − (4 pˆ 11 ( x, z ) − 1) ] fˆY (Mˆ Y )

aˆ 2 =

(4 pˆ 11 ( x, z ) − 1)[(4 pˆ 11 ( x, y ) − 1) − (4 pˆ 11 ( y, z ) − 1)(4 pˆ 11 ( x, z ) − 1)] Mˆ Z fˆZ (Mˆ Z ) ⋅ [1 − (4 pˆ 11 (x, z ) − 1)2 ] fˆY (Mˆ Y )

a3 =

[(4 pˆ 11 ( y, z ) − 1) − (4 pˆ 11 (x, y ) − 1)(4 pˆ 11 (x, z ) − 1)] Mˆ Z fˆZ (Mˆ Z ) ⋅ [1 − (4 pˆ 11 (x, z ) − 1)2 ] fˆY (Mˆ Y ) (6.9)

are the sample estimates of a1, a2 and a3 given in (6.5) respectively, F*(⋅) is a function of

(Mˆ

)

, u , v, w, aˆ1 , aˆ 2 , aˆ 3 such that

F * (T *) = M Y ⇒ F1 * (T *) =

∂F * (⋅) ∂Mˆ Y

=1 T*

F2 * (T *) =

∂F * (⋅) = − a1 ∂u T *

F3 * (T *) =

∂F * (⋅) = −a 2 ∂v T *

F4 * (T *) =

∂F * (⋅) = −a3 ∂w T *

F5 * (T *) =

∂F * (⋅) =0 ∂aˆ1 T *

F6 * (T *) =

∂F * (⋅) =0 ∂aˆ 2 T *

F7 * (T *) =

∂F * (⋅) =0 ∂aˆ 3 T *

where T* = (MY,1,1,1,a1,a2,a3) Under these conditions it can easily be shown that

(

)

( )

( F *) E Mˆ Y = M Y + 0 n −1

ˆ ( F *) is given by and to the first degree of approximation, the variance of M Y

( ) ( ˆ ( ) ) is given in (6.6). where min.Var (M

F ( F *) = min.Var Mˆ Y Var Mˆ Y

)

(6.10)

Under the cost function (5.8), the optimum values of m and n which minimizes the minimum variance of (F ) Mˆ Y is (6.6) are given by

m opt =

(V0 − V1 − V3 ) / C1 (V0 − V1 − V3 )C1 + (V1 − V2 − V3 )(C 2 + C3 )] C0

[

(6.11)

(V1 − V2 − V3 ) / C 2 (V0 − V1 − V3 )C1 + (V1 − V2 + V3 )(C 2 + C3 )] C0

n opt =

[

where

D 2V0

V3 =

[1 − (4P (x, z ) − 1) ] ˆ ( ) ) is given by for large N, the optimum value of min.Var (M ( ) )] = [ (V − V − V )C + C(V − V Opt.[min.Var(Mˆ

(6.12)

+ V3 )(C 2 + C 3 )

]

(6.13)

The proposed two-phase sampling strategy would be profitable over single phase-sampling so long as

[ ( )]

[

(

ˆ (F ) Opt. Var Mˆ Y > Opt. min.Var M Y

)]

C 2 + C 3  V0 − V0 − V1 − V3  i.e. <  c1 V1 − V2 + V3  

(6.14)

It follows from (5.7) and (6.13) that

[

(

Opt. min.Var Mˆ Y if

 V0 − V1 − V0 − V1 − V3   V1 − V2 + V3 

(F )

)] < Opt.[min.Var(Mˆ )] H

  C 2 + C3 > −   C1  

for large N. Further we note from (5.11) and (6.13) that

V1 C2  (V1 − V2 +V 3 )C1 C1 

(6.15)

[

(

)]

[

(

G (F ) (g ) Opt. min.Var Mˆ Y < Opt. min.Var Mˆ Y orMˆ Y

C 2 + C 3  (V0 − V1 ) − (V0 − V1 − V3 )  <  C1  (V1 − V2 + V3 ) − V1 − V2 

)]

(6.16)

REFERENCES Chand, L. (1975): Some ratio-type estimators based on two or more auxiliary variables. Unpublished Ph.D. dissertation, Iowa State University, Ames, Iowa. Francisco, C.A. and Fuller, W.A. (1991): Quntile estimation with a complex survey design. Ann. Statist. 19, 454-469. Kiregyera, B. (1980): A chain ratio-type estimator in finite population double sampling using two auxiliary variables. Metrika, 27, 217-223. Kiregyera, B. (1984): Regression-type estimators using two auxiliary variables and the model of double sampling from finite populations. Metrika, 31, 215-226. Kuk, Y.C.A. and Mak, T.K. (1989): Median estimation in the presence of auxiliary information. J.R. Statist. Soc. B, (2), 261-269. Sahoo, J. and Sahoo, L.N. (1993): A class of estimators in two-phase sampling using two auxiliary variables. Jour. Ind. Statist. Assoc., 31, 107-114. Singh, S., Joarder, A.H. and Tracy, D.S. (2001): Median estimation using double sampling. Aust. N.Z. J. Statist. 43(1), 33-46. Singh, H.P. (1993): A chain ratio-cum-difference estimator using two auxiliary variates in double sampling. Journal of Raishankar University, 6, (B) (Science), 79-83. Srivenkataramana, T. and Tracy, D.S. (1989): Two-phase sampling for selection with probability proportional to size in sample surveys. Biometrika, 76, 818-821. Srivastava, S.K. (1971): A generalized estimator for the mean of a finite population using multiauxiliary information. Jour. Amer. Statist. Assoc. 66, 404-407. Srivastava, S.K. and Jhajj, H.S. (1983): A class of estimators of the population mean using multi-auxiliary information. Cal. Statist. Assoc. Bull., 32, 47-56.