The GNSS integer ambiguities: estimation and validation
The GNSS integer ambiguities: estimation and validation
The GNSS integer ambiguities: estimation and validation
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where y ∈ R m with mean E{y} = Aa + Bb <strong>and</strong> PDF fy(y), <strong>and</strong> ˜ θ is the BIE estimator<br />
of θ = l T a a + l T b b.<br />
<strong>The</strong> BIE estimator can also be written as:<br />
˜θ = l T a ã + l T b ˜ b (4.5)<br />
with<br />
⎧<br />
⎪⎨ ã =<br />
⎪⎩<br />
<br />
z∈Zn zwz(y), <br />
z∈Zn wz(y) = 1<br />
<br />
˜b = ζwζ(y)dζ,<br />
<br />
wζ(y)dζ = 1<br />
R p<br />
R p<br />
(4.6)<br />
where the weighting functions wz(y) <strong>and</strong> wζ(y) follow from equation (4.4). This shows<br />
that the BIE estimator of the <strong>integer</strong> parameter vector a is also a weighted sum of all<br />
<strong>integer</strong> vectors in Zn , just like the admissible <strong>integer</strong> estimators of (3.6). However, in<br />
this case the weights are not binary (1 or 0); their values are determined by y <strong>and</strong> its<br />
PDF, so that the BIE weights are real-valued <strong>and</strong> nonzero for all <strong>integer</strong> vectors. This<br />
implies that the estimator ã is real-valued, instead of <strong>integer</strong>-valued. Furthermore, it<br />
implies that there is no need to apply a discrimination test.<br />
In Teunissen (2003f) it was shown that the BIE estimator is unbiased <strong>and</strong> has minimum<br />
variance, i.e. it has better precision than the best linear unbiased estimator. This result<br />
applied to the baseline estimator gives:<br />
⎧<br />
⎪⎨ D{<br />
⎪⎩<br />
˜b} ≤ D{ ˇb} <strong>and</strong> E{ ˜b} = E{ ˇb} = b<br />
D{ ˜b} ≤ D{ ˆb} <strong>and</strong> E{ ˜b} = E{ ˆ (4.7)<br />
b} = b<br />
Hence, the BIE baseline precision is always better than or as good as the precision of its<br />
float <strong>and</strong> fixed counterparts.<br />
In <strong>GNSS</strong> applications it is assumed that the data are normally distributed as mentioned<br />
in section 3.1.3. It can be shown that in this case the BIE estimator is given by equation<br />
(4.5) with the following estimators for the <strong>ambiguities</strong> <strong>and</strong> the baseline parameters:<br />
⎧ P<br />
⎪⎨ ã =<br />
⎪⎩<br />
z∈Zn z exp{− 1<br />
2 â−z2Q<br />
}<br />
â<br />
P<br />
exp{− 1<br />
2 â−z2 Q }<br />
â<br />
z∈Z n<br />
˜ b = ˆ b − Qˆ bâ Q −1<br />
â<br />
See appendix C.<br />
(â − ã)<br />
<br />
= zwz(â)<br />
z∈Z n<br />
(4.8)<br />
Note that the formal expressions in (4.8) are identical to their Bayesian counterparts as<br />
presented in section 3.6, but that the distributional properties of the BIE estimator <strong>and</strong><br />
its Bayesian counterpart of course differ. Since the functional form of the non-Bayesian<br />
estimator is identical to the Bayesian solution in the special case of normally distributed<br />
data, the theory of BIE <strong>estimation</strong> has provided the link with the Bayesian approach of<br />
ambiguity resolution.<br />
<strong>The</strong> BIE estimator 69